09.03.2021  Ausgabe 2/2021 Open Access
Nonrandom sampling and association tests on realized returns and risk proxies
 Zeitschrift:
 Review of Accounting Studies > Ausgabe 2/2021
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1 Introduction
In this paper, we develop, validate, and illustrate a practicable approach to deal with nonrandom samples whose underlying cause is data requirements, a pervasive issue in accounting research. Examples include requirements for analyst following, database inclusion (e.g., Execucomp includes S&P 1500 firms), and stock price above a threshold, such as $5. We examine how nonrandomness
^{1} of the dependent variable in datarestricted samples affects results of empirical association tests, propose and validate a nonparametric resampling technique (“distributionmatching”) to adjust for the effects of nonrandomness, and thereby increase the generalizability of results and apply the technique to tests of associations between realized returns and implied cost of equity
(CofE) metrics. Our goal is to assist accounting researchers in constructing more powerful and less biased test samples, thereby increasing the generalizability (decreasing the sampledependency) of results.
As discussed later, distributionmatching differs from selection models and multiple imputations, which are sometimes used in the analysis of datarestricted samples. A Heckmantype selection model approach assumes the selection model can be reliably estimated on a random sample of the population, but practical research settings typically involve a tradeoff between selection model fit and data requirements for the selection model variables. A selection model approach may therefore simply transfer the datarestrictions issue from the test model to the selection model. In our setting, we find that introducing a selection model does not resolve the issue of nonrandomness that we address using distribution matching; in contrast, applying multiple imputation yields results that converge to the distributionmatching results.
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The starting point of the distributionmatching technique is a common requirement in archival accounting research that the sample contains only observations with complete data for all variables of interest (“complete cases”). We first examine a pairwise association in a stylized simulation setting consisting of a reference sample with full information on one variable (
y) and restricted data on the second variable (
x). Completecase analyses effectively impose any data restrictions of
x on
y and do not use information about
y in incomplete observation pairs. If missingness of
x is even weakly correlated with values of
y, completecase samples are nonrandom samples of
y and associationtest results would not generalize to the reference sample. Distributionmatching uses information about the marginal distribution of
y (the reference distribution) and resamples completepairs of observations (
x and
y) from the datarestricted sample to match the reference distribution as closely as possible. The goal is to construct samples that appear as if they were drawn randomly from the reference distribution, despite the data restrictions, and using only the observations of the datarestricted sample. Using simulated data with known induced levels of statistical associations and three types of nonrandomness, we show that association tests yield biased results in the nonrandom samples; applying distributionmatching substantially reduces or even eliminates the bias. We view the results of these simulation analyses as providing evidence that distributionmatching can address the issue of nonrandom sampling in a generic research setting.
To illustrate distributionmatching in a specific research setting, we apply the technique to an archival setting characterized by stringent data requirements and inconsistent/counterintuitive results (e.g., Easton and Monahan
2005), namely, tests of the association between realized returns and
CofE metrics. By definition, all firms
have a cost of equity, but a researcher cannot
observe the cost of equity for some firms, typically because of data requirements.
^{2} The theory linking realized returns and
CofE metrics applies to the reference sample defined in note 1 (CRSP firms with at least 12 months of realized returns during February 1976–July 2009). This reference sample contains complete data on one variable (
y, returns) and not on the other variable (
x,
CofE).
^{3} We show (1) the unadjusted sample with data on
CofE metrics is a nonrandom sample of the reference sample (the
CofE sample returns have substantially lower standard deviation, skewness and kurtosis)
^{4}; and (2) tests of the
CofEreturns association based on the unadjusted
CofE sample—that is, a sample that would typically be used in accounting research—produce weak or negative associations between realized returns and implied cost of equity metrics, consistent with previous empirical research (e.g., Easton and Monahan
2005) and inconsistent with theory.
In contrast, tests using a distributionmatched
CofE sample produce statistically reliable evidence of the theoreticallypredicted positive association between
CofE metrics and realized returns. Distributionmatching does not involve creating new data; the technique resamples
CofEsample observations (
CofE and returns pairs) so that the returns distribution in the distributionmatched
CofE sample mimics the returns distribution in the reference sample. We apply two approaches. The first uses the nonparametric KolmogorovSmirnov (KS) test of general sample differences. The KS statistic rejects, at the 0.10 (0.05) [0.01] level, the hypothesis of distribution equality in the unadjusted
CofE sample, compared to the reference sample in 401 (401) [393] of the 402 sample months. We distributionmatch by constructing monthly subsamples using only
CofEsample observations (
CofE and returns pairs) so that the deviation between the returns distribution of the resulting distributionmatched
CofE sample and the reference distribution of returns, as captured by the KS statistic, is minimized. This resampling procedure is effective and requires few assumptions but imposes a substantial computational burden. To address this practical concern, the second approach sorts the returns distributions of both the reference sample and the
CofE sample into researcherdefined strata (“bins”) of the continuous variable and applies a form of stratified resampling that aims to match the standard deviation of the reference returns distribution. We find that distributionmatching can result in smaller samples than the original nonrandom samples; despite a possible loss of power, correlations between realized returns and
CofE metrics are with one exception reliably positive in distributionmatched samples obtained using both approaches.
^{5}
The methodological inference from our results is that selection criteria yielding samples with outcome distributions differing from the reference distribution can materially affect the results of association tests, including producing results that do not generalize to the reference sample to which the tested hypothesis applies. We extend this inference in several ways. To illustrate the inference in other settings with restrictive data requirements, we show that imposing several plausible selection criteria on the reference sample (S&P 500 membership, NYSE listing, availability of a dispersion measure of analyst earnings forecasts, and stock price at least $5) can change the distribution of realized returns and lead to biased estimates in association tests of realized returns with risk factor premia. We reach the same inference when we directly induce changes in the distribution of realized returns and show the sensitivity of risk factor premia to these changes. To illustrate that applying distributionmatching does not produce false results, we apply the technique to Richardson et al.’s (
2005) analysis of the association between returns and accruals, a setting in which previous research shows results consistent with theory, and do not overturn their inferences. To separate the effects of reduced sample size from the effects of nonrandomness, we compare the association between realized returns and assetpricing factor betas for a random subsample from the full returns sample (to capture the effects of reduced sample size) and the actual
CofE sample (to capture the combined effects of reduced sample size and nonrandomness). We reason that imposing an unnecessary data restriction on samples used in an association test with risk metrics (factor betas) that can be performed on the entire reference sample allows us to separate the effects of nonrandomness from the effects of a reduced sample size. The coefficients on factor betas
^{6} for the random sample of equal size as the
CofE sample are similar in magnitude and statistical significance to those for the reference sample, while for the actual
CofE sample there is no reliable association between realized returns and any factor beta. These results suggest that, (1) in our setting, efficiency losses due to reduced sample sizes alone have little effect on qualitative inferences and (2) the
CofE sample should not be assumed to be a random subsample of the reference sample. Finally, we show that inferences from analysis of our main
CofE sample, restricted by the data requirements of all five
CofE metrics we consider, are qualitatively similar when we redo our main association tests on a purely IBESbased
CofE sample.
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We believe our findings support a conclusion that results obtained using unadjusted nonrandom samples may not support generalizations to a researcherselected reference sample. In fact, our analysis of the
CofE sample highlights that maximizing the size of the nonrandom sample, after imposing data requirements, may conflict with the goal of obtaining a random sample, which is fundamental for the generalizability of results. We also believe our analyses provide a practical solution to this issue, in the form of distributionmatching.
2 Motivation for and validation of distributionmatching
2.1 Motivation and intuition
In accounting research settings, data constraints often mean that analyses can be performed only on a subsample of observations, even though the results are intended to generalize to a population or reference sample to which the tested hypothesis logically applies.
^{7} The setting we consider is association tests (regression coefficients or correlation coefficients) between two variables, of which one (
y) is available for all firms in the researcherdefined reference sample, while the second variable (
x) is often missing.
^{8} A common treatment in the accounting literature is restricting the test sample to observations with complete information, yielding a datarestricted sample. The data constraints on
x are imposed on
y, causing information about the unrestricted distribution of
y to be lost. We hypothesize (1) this deletion leads to nonrandom test samples and (2) association tests using these samples yield results that may not be generalizable to the reference sample. We address this problem by incorporating information about the reference distribution (of the complete variable
y) into the association test. We resample observations from the datarestricted nonrandom sample to create an adjusted sample that mimics the reference distribution of the complete variable and appears randomly drawn from the reference sample with respect to
y, despite data constraints on
x.
The intuition for this approach is as follows. Consider the estimate of a Pearson correlation coefficient
\( \hat{\rho} \) between two continuous random variables
x and
y
^{9}:
where
x
_{i} and
y
_{i} are standardized (demeaned and divided by their respective standard deviations) realizations of
x and
y,
f(.) denotes the density function, and
s
_{i} is an observationlevel indicator for membership in the datarestricted test sample. For simplicity, subscripts for time
t are suppressed.
$$ \hat{\rho}=\int \int {x}_i\;{y}_i\;f\left({x}_i,{y}_i{s}_i=1\right)\; dxdy=\int \int {x}_i\;{y}_i\;f\left({x}_i{y}_i,{s}_i=1\right)\;f\left({y}_i{s}_i=1\right)\; dxdy, $$
(1)
The true correlation in the reference sample, assuming availability of complete data, is given by:
$$ {\rho}^{\ast }=\int \int {x}_i\kern0.28em {y}_i\kern0.28em f\left({x}_i,{y}_i\right)\kern0.28em dxdy=\int \int {x}_i\kern0.28em {y}_i\kern0.28em f\left({x}_i{y}_i\right)\kern0.28em f\left({y}_i\right)\kern0.28em dxdy. $$
(2)
\( \hat{\rho} \) is a consistent estimator of the true
ρ
^{∗} only if the joint distribution in the restricted sample equals the joint distribution in the reference sample,
f(
x
_{i},
y
_{i}
s
_{i} = 1) =
f(
x
_{i},
y
_{i}) or, equivalently,
f(
x
_{i}
y
_{i},
s
_{i} = 1)
f(
y
_{i}
s
_{i} = 1) =
f(
x
_{i}
y
_{i})
f(
y
_{i}). This condition implies that the unobserved data are missing completely at random (MCAR); only then would a restricted sample (i.e., a sample after deletion of observations because of missing data) be a random subsample of the reference sample.
In our stylized setting, as well as in some other accounting research settings, it is possible to assess the difference in the marginal distributions of the fully observed variable
y
_{i} between the restricted sample,
f(
y
_{i}
s
_{i} = 1), and the reference sample,
f(
y
_{i}) and reject the assumption of MCAR. Differences between these marginal distributions mean the restricted sample is nonrandom and consistency of
\( \hat{\rho} \) is less likely.
If the MCAR assumption is rejected, it must be replaced with a weaker assumption: either the data are missing at random, conditional on observed variables (MAR), including
y (realized return in our application), or the data are not missing at random (NMAR), which implies that missingness also depends on unobserved data. While it is possible to reject the MCAR condition, the unavailability of missing data precludes testing whether data are MAR or NMAR. Our main analyses extend the common approach of constructing test samples by listwise deletion under the MCAR assumption. We begin by showing differences in the marginal distributions of realized returns between a fullreturns sample and a
CofE subsample and that results of association tests (with factor betas) also differ qualitatively between these two samples. We therefore focus on methods under the MAR assumption, specifically, distributionmatching and multiple imputations. In Section
5, we assess the impact of a possible NMAR assumption using a Heckmantype selection model.
Referring to Eqs. (
1) and (
2), the distributionmatching approach requires the distribution of
x, conditional on
y
_{i}, is unchanged in the datarestricted sample, compared to the reference sample:
$$ f\left({x}_i{y}_i,{s}_i=1\right)=f\left({x}_i{y}_i\right). $$
(3)
Assuming complete data on
y, the MAR assumption implies Eq. (
3) holds.
^{10} Then
\( \hat{\rho} \) will converge to
ρ
^{∗} as
f(
y
_{i},
s
_{i} = 1) approaches
f(
y
_{i}) via distributionmatching. In the context of our archival analysis, condition (
3) implies the
CofE metrics are not systematically biased in the restricted sample, conditional on the value of the future realized return. It seems unlikely that the probability of having the analysts’ forecasts required to construct an implied
CofE metric depends on the value of realized returns, which can be assessed only ex post.
Regardless of concerns specific to our application, condition (
3) contrasts with and is arguably weaker than the assumption that data are missing at random, essentially equating
f(
x
_{i},
y
_{i}
s
_{i} = 1) with
f(
x
_{i},
y
_{i}), particularly when differences in the marginal distributions of returns between the reference sample and the
CofErestricted sample,
f(
y
_{i}) versus
f(
y
_{i}
s
_{i} = 1), are knowable from the data. We use the marginal distribution of the dependent variable of the reference sample as opposed to simply deleting observations with incomplete data. That is, under condition (
3), our approach focuses on the marginal distribution of
y in the datarestricted and possibly nonrandom sample. We resample systematically only from observations in this sample with complete data on both variables, so that, in the limit, the marginal distribution of
y
_{i} matches the marginal reference distribution of
y
_{i}:
$$ f\left({y}_i{s}_i=1\right)\to f\left({y}_i\right). $$
(4)
While the convergence in (4) is achievable in the limit, the effectiveness of distributionmatching in a given research setting is a function of, among other things, the number of restrictedsample observations and the size of the common support of the restrictedsample and reference distributions. A smaller restricted sample means fewer observations to resample from and a smaller common support of the distributions means
f(
y
_{i}) is more severely truncated in the restricted sample. In addition, the resampling approach may be unnecessary if only a few observations are missing, making the restricted sample (nearly) equal to the reference sample.
To measure similarity in the cumulative distributions between the reference sample and the nonrandom datarestricted sample, we use the nonparametric KolmogorovSmirnov (KS) statistic, which computes the percentage maximum absolute distance between two cumulative empirical distributions.
where
F
^{NRS}(
y
_{i}),
F
^{POP}(
y
_{i}) are the cumulative distributions of
y in the nonrandom sample and population, respectively. We use the KS statistic and its associated asymptotic
p value for a test of distribution equality between a subsample and the reference sample and to assess the degree of convergence in (4) within the KSbased distributionmatching approach. Information on the setup of the simulation is available from the corresponding author.
$$ KS=\underset{i}{\max}\left{F}^{NRS}\left({y}_i\right){F}^{POP}\left({y}_i\right)\right\kern0.24em \;\;\;\mathrm{where}\ i=1,2,\dots, n, $$
(5)
2.2 Relation of distributionmatching to traditional matching approaches and other missingdata approaches
Traditional matching approaches The distributionmatching approach differs substantially in goal and implementation from traditional matching techniques that might be used to address issues of endogeneity and sample selection on observable determinants. Traditional matching relies on an observationbyobservation comparison (for example, matched pairs), while distributionmatching aims to reweight the observations in a sample distribution so that the resulting distribution approximates an (empirical or theoretical) reference distribution. Distributionmatching focuses on the outcome variable, while traditional matching focuses on (possibly multiple) independent variables, for example, through sorting or propensity scoring observations.
Approaches applicable in a MAR setting
Missingdata approaches under the MAR assumption include multiple imputation (MI) and fullinformation maximum likelihood (FIML) estimation.
^{11} FIML incorporates information about the marginal distribution
f(
y
_{i}) by including the observations in the likelihood calculation, even if data on some variables are missing. MI uses a stochastic regression framework to impute possible values for the missing data multiple times, after which the completed (“imputed”) datasets can be independently analyzed and the results aggregated. Complete variables, that is, the marginal distribution of returns in our setting, are preserved from the reference sample and considered in the analysis. MI uses the entire reference sample, so it is more efficient than distributionmatching in crosssectional analyses; it can be applied when data are missing for more than one variable; and it can incorporate the use of auxiliary variables that are either informative about missingness or correlated with the missing data.
^{12} However, distributionmatching is nonparametric while both MI and FIML rely on multivariate normality. Descriptive statistics in Table
2 suggest the normality assumption is unlikely to hold in our example setting with realized returns. Based on theoretical arguments in Schafer (
1997), supported by simulation evidence of Demirtas et al. (
2008), that MI appears to be less susceptible to deviations from multivariate normality than maximum likelihood, we repeat our main tests using subsamplesbased forms of MI and find results similar to those obtained using distributionmatching.
Other missingdata approaches
We clarify the intuition of distributionmatching by contrasting it with three other approaches: estimations that take account of truncation; incomplete poststratification and selection models. First, assuming the true distribution of the complete outcomes is normal, Tobin (
1956) derives closedform solutions when the outcome variable is truncated at a known upper or lower bound (see also Wooldridge
2010). Rather than focus on truncation, we emphasize that nonrandomness likely manifests in a restricted sample with a different shape than the reference distribution, even if the common support is large or complete.
^{13} Also, rather than making assumptions about the reference distribution of the outcome variable, we estimate its shape from the reference sample with complete data.
^{14} Second, the survey literature uses incomplete poststratification, which involves reweighting observations according to their marginal weights in a reference distribution or population. The weights are typically constructed based on discrete and exogenous variables, such as gender, not an outcome variable. The similarity to distributionmatching arises because survey respondents need not be representative of the population, necessitating reweighting responses if the goal is to generalize results to the population.
^{15} To that end, both poststratification and distributionmatching import information about the marginal reference distribution. In fact, for the common support region of the sample and reference distributions, distributionmatching is essentially a form of poststratification that treats the variable as continuous (each
y
_{i} is its own stratum) and does not require exante grouping of observations into strata. In addition, the sample distribution may not only be nonrandom within the common support region but also truncated when compared to the reference distribution. Intuitively, the effect of truncation is mitigated by oversampling from the tails of the sample distribution. In short, distributionmatching tries to combine the notions of stratified sampling and overcoming biases from truncation.
Third, distributionmatching assumes data are missing at random, conditional on observables, while Heckmantype selection models assume data are
not missing at random, necessitating a firststage probit selection model to capture the mechanism that selects observations into the restricted sample. Under certain conditions,
^{16} the bias in the test model can be alleviated by incorporating the inverse Mills ratio from the selection model. In contrast, distributionmatching disregards the selectionmechanism and uses information about its consequences by assessing and minimizing the difference of the sample distribution to a reference distribution. In Section
5.1, we apply the selection model approach and find that results using the restricted nonrandom samples, both on
CofE and on factor betas, are little affected by including the inverse Mills ratio.
2.3 Validity tests on simulated data
We use simulated data to validate our distributionmatching approach by showing that correlation estimates from distributionmatched samples converge to their true values, even though these samples consist
only of nonrandomly drawn observations from the reference sample.
^{17} We generate populations of data for two variables (
y and
x) with known correlations and draw from these simulated populations both randomly and in three nonrandom ways, with selection probabilities based on the marginal distribution of
y. For each of the three types of nonrandom samples drawn from the simulated populations, we resample with replacement to create distributionmatched samples. Using
only observations from the respective nonrandom sample, distributionmatching is designed to mimic the marginal distribution of variable
y in the population as closely as possible.
The outcome variable of interest is the estimated correlation between
y
_{i} and
x
_{i} in the nonrandom samples before and after distributionmatching. We examine population correlations specified at 0.5, −0.5, and 0 (to rule out the possibility that distributionmatching induces a correlation where none exists). We examine both negative and positive true correlations to provide evidence that the effectiveness of distributionmatching does not depend on the either the sign of the true association or the sign of the bias in the association estimate. We draw three kinds of nonrandom samples. In Nonrandom sample I, the selection probability of a given observation is decreasing in the absolute distance from the mean, leading to fewer observations that include extreme
y values. Nonrandom sample II samples observations based on the uniform distribution over the entire interval of
y observations, leading to higher selection probabilities for observations in the tails. These two symmetric nonrandom samples are expected to yield either a negative bias (Nonrandom sample I) or a positive bias (Nonrandom sample II) in the estimates of associations. Nonrandom sample III is a form of nonsymmetric selection probability, in that selection probability of an observation is increasing in
y.
Figure
1 presents visual evidence of the effectiveness of distributionmatching for the three types of nonrandom samples, drawn from normally distributed population data. The figure depicts example distributions of
y for a single randomlychosen simulation run, for the unadjusted nonrandom samples (on the left), and the distributionmatched samples (on the right) of size
m = 1000. The benchmark distribution, which appears in both right and left graphs, is from the population in that particular run (
n = 5000). For all three nonrandom draws, the leftside graphs illustrate that the distributions deviate from the population benchmark. After distributionmatching, the rightside graphs show the sample distributions closely follow the population distribution and are indistinguishable for large regions of
y.
×
Table
1 reports numerical results of the simulations analysis. We focus on the results in Panel A (normally distributed variables). Results in Panel B (nonnormally distributed variables) are qualitatively similar, suggesting the effectiveness of the nonparametric distributionmatching approach does not depend on the shape of the marginal distributions, in particular, a normality assumption. We first verify that empirical correlation estimates in the population and random samples (CORR) are close to the specified (true) correlations (CORR*), and there are no meaningful differences between the population and the random sample. For all three levels of true correlation, the KS statistic for random samples is about 2.4%, and
p values for the difference between the random sample and the population are about 0.69. Estimated correlations differ from true correlations by 0.005 or less, confirming that a reduction in sample size, even to 20% of the population (
m = 1000), is unimportant for the associationtest point estimates as long as the sample selection is random.
Table 1
Simulation results from forced nonrandom samples and corresponding distributionmatched samples
CORR*=0.5

CORR*=0

CORR*=−0.5



KS

CORR

KS

CORR

KS

CORR


Panel A: Both variables (standard) normally distributed


Population (n = 5000)

N/A

0.5000

N/A

0.0003

N/A

−0.4999

Random Sample (m = 1000)

0.0244

0.4991

0.0245

−0.0001

0.0243

−0.5005

(p value)

(0.6904)

(0.6878)

(0.6944)


NonRandom Sample I


Ranked abs. distance to mean

0.1385

0.3267

0.1403

0.0006

0.1403

−0.3268

(p value)  %ile (Random Sample)

(0.0000)

0.0

(0.0000)

50.5

(0.0000)

100.0

DistributionMatched Sample

0.0302

0.4856

0.0297

0.0018

0.0299

−0.4877

(p value)  %ile (Random Sample)

(0.5422)

30.1

(0.5486)

53.6

(0.5391)

71.4

NonRandom Sample II


Uniform distribution

0.2550

0.7775

0.2546

0.0015

0.2533

−0.7780

(p value)  %ile (Random Sample)

(0.0000)

100.0

(0.0000)

53.2

(0.0000)

0.0

DistributionMatched Sample

0.0093

0.4953

0.0095

0.0014

0.0094

−0.4943

(p value)  %ile (Random Sample)

(0.9992)

43.7

(0.9992)

53.2

(0.9987)

60.7

NonRandom Sample III


Ranked abs. distance to maximum

0.2665

0.4236

0.2628

0.0017

0.2645

−0.4233

(p value)  %ile (Random Sample)

(0.0000)

0.0

(0.0000)

53.5

(0.0000)

99.9

DistributionMatched Sample

0.0348

0.4873

0.0348

0.0034

0.0347

−0.4831

(p value)  %ile (Random Sample)

(0.4003)

31.5

(0.4078)

55.0

(0.4073)

77.5

Panel B: Both variables nonnormally distributed


Population (n = 5000)

N/A

0.4999

N/A

0.0005

N/A

−0.5005

Random Sample (m = 1000)

0.0245

0.5010

0.0246

0.0009

0.0244

−0.5001

(p value)

(0.6851)

(0.6833)

(0.6921)


NonRandom Sample I


Ranked abs. distance to mean

0.1419

0.3704

0.1450

−0.0003

0.1428

−0.3422

(p value)  %ile (Random Sample)

(0.0000)

0.0

(0.0000)

49.2

(0.0000)

100.0

DistributionMatched Sample

0.0326

0.5018

0.0332

0.0003

0.0323

−0.4879

(p value)  %ile (Random Sample)

(0.4691)

50.6

(0.4328)

48.9

(0.4588)

67.8

NonRandom Sample II


Uniform distribution

0.4993

0.7728

0.5033

0.0024

0.4976

−0.7500

(p value)  %ile (Random Sample)

(0.0000)

100.0

(0.0000)

51.2

(0.0000)

0.0

DistributionMatched Sample

0.0135

0.5023

0.0135

0.0007

0.0134

−0.4993

(p value)  %ile (Random Sample)

(0.9837)

51.4

(0.9831)

49.0

(0.9846)

52.0

NonRandom Sample III


Ranked abs. distance to maximum

0.2649

0.4305

0.2615

0.0007

0.2654

−0.4253

(p value)  %ile (Random Sample)

(0.0000)

0.1

(0.0000)

49.0

(0.0000)

99.6

DistributionMatched Sample

0.0355

0.4808

0.0349

−0.0002

0.0355

−0.4887

(p value)  %ile (Random Sample)

(0.4070)

17.7

(0.3933)

48.1

(0.4034)

67.1

In contrast, and by construction, nonrandom samples have a distribution of
y that differs sharply from the population distribution. For Nonrandom sample I (Nonrandom sample II) [Nonrandom sample III], KS statistics are about 13.9% (25.5%) [26.7%]. To assess the significance of the bias in correlation estimates for these samples, we report the percentile of the mean nonrandom sample correlation in the distribution spanned by the 1000 correlations as ‘Percentile (Random Sample).’ A small (large) percentile corresponds to a low (high) estimate. When the true correlations are 0.5 or −0.5, the estimated correlations are biased towards zero in magnitude in Nonrandom samples I and III and are upward biased in Nonrandom sample II. The bias is highly significant, with percentiles of either 0.0 (i.e., below the distribution of 1000 correlations from random samples) or 99.9 or higher (i.e., only one or fewer of the 1000 correlations is higher). Distributionmatching reduces or eliminates the effects of nonrandom sampling: Across all three nonrandom samples, the corresponding distributionmatched samples exhibit correlations much closer to the true correlations. Biases are close to zero (never exceeding 0.0169 in magnitude) and are insignificant in all cases, with percentiles ranging from 30.1 to 77.5.
Based on these simulation results, we conclude that distributionmatching is effective in reducing the bias in correlation estimates in nonrandom samples of
y
_{i}; results for zero correlations show that distributionmatching does not induce an (apparent) correlation where none exists. The result for Nonrandom sample I is of particular interest. Because the result is based on simulated data and an imposed criterion in the sample construction, we view this finding as suggesting the kind of bias in association tests if data availability requirements bias a sample toward firms with typicallylessextreme returns realizations (described by prior research as relatively “stable” firms; see Footnote 4), as is often the case in accounting research situations. Our next tests investigate whether this simulation result applies to a specific wellstudied empiricalarchival setting.
3 Application to the association of realized returns and implied cost of equity metrics
We test for a bias in association test results when samples are restricted to firms with available data to calculate analystbased implied
CofE metrics. Specifically, we reexamine the correlation between realized returns and
CofE metrics, as they have been used to test for expectedreturns associations. Settings include voluntary disclosure (Botosan
1997), AIMR scores (Botosan and Plumlee
2002), earnings attributes (Francis et al.
2004), restatements (Hribar and Jenkins
2004), legal institutions/security regulation (Hail and Leuz
2006), shareholder taxes (Dhaliwal et al.
2007), mandatory IFRS adoption (Li
2010), earnings quality and information asymmetry (Bhattacharya et al.
2012), and financial constraints and taxes (Dai et al.
2013). We believe the construct validity of analystbased
CofE metrics is of interest to many researchers, so that an application of the distributionmatching approach in this setting provides insights in its own right. Section
3.1 summarizes previous research, and Section
3.2 explains why we chose this setting to illustrate distributionmatching.
3.1 Research on the association between realized returns and implied CofE metrics
Realized returns are the dependent variable in a variety of association tests, including twostage crosssectional asset pricing tests (associations of realized returns with risk factor betas, Fama and MacBeth
1973) and cost of equity tests (associations of realized returns with implied
CofE metrics). The latter are predicated on the view that both realized returns and
CofE metrics are potentially noisy or confounded proxies for unobservable expected returns. Intuitively, a firm’s expected return should be commensurate with its riskiness. Realized returns are expost outcome measures, affected by the arrival of information during the return measurement period and therefore contain an expected return component and a potentially nonzero unexpected return component caused by news about cash flows and news about the expected return itself (Campbell and Shiller
1988; Campbell
1991).
Researchers typically assume
either that (1) realized returns are a reasonable proxy for expected returns (that is, the unexpected return component is small, cancels out through aggregation, or both)
or, (2) even in broad samples, the unexpected return component is a key noncancelling component of realized returns (e.g., Elton
1999; Vuolteenaho
2002).
^{18} Adopting the latter perspective, researchers have developed several
CofE metrics derived independently of realized returns (e.g., Gebhardt et al.
2001; Claus and Thomas
2001; Botosan and Plumlee
2002; Easton
2004; Brav et al.
2005; Ohlson and JüttnerNauroth
2005),
^{19} or, alternatively, researchers have empirically purged the realized return of its unexpected (“news”) component. For example, Campbell (
1991) and Vuolteenaho (
2002) propose a variance decomposition method that prespecifies expected returns as a linear combination of firm characteristics; Botosan et al. (
2011) and Ogneva (
2012) control for a specific kind of fundamental (earnings) news in realized returns to identify the cashflow news component of realized returns.
^{20} A stream of research that views realized returns and
CofE metrics as alternative and imperfect proxies for expected returns aims to validate
CofE metrics jointly with realized returns in association tests. Easton and Monahan (
2005) and Guay et al. (
2011) find the association between realized returns and several commonly used
CofE estimates is often insignificant or even significantly negative. Botosan et al. (
2011) find the association varies between positive and negative over time and is, on average, weak.
The contrast of these results with economic intuition leads some researchers to propose and test approaches to increase the association. One approach attributes the weak association to realized returns. Using variance decomposition to control for nonexpected return components in realized returns, Easton and Monahan (
2005) find no, or significantly negative, associations between “newspurged” realized returns and four of the seven
CofE estimates they consider. In contrast, Botosan et al. (
2011) use different empirical proxies for unexpected return components (similar to Ogneva
2012) and find their
CofE estimates have significant positive associations with “newspurged” returns. However, they also document their newspurged returns measure has either no association or counterintuitive associations with the riskfree rate, beta, booktomarket, and other proxies for risk, leading them to question the validity of their “newspurged” realized returns as a proxy for expected returns and to express caution about the approach. In terms of our research setting, we note there seems to be a tradeoff in that adjustments to
CofE metrics may worsen the relation between
CofE metrics and other proxies for risk, such as beta (Botosan et al.
2011). We address this potential concern in Section
4.4.2 by showing the coefficients on risk factors are little affected by our distributionmatching approach.
Other papers attribute the weak association to the analyst forecastinputs. Guay et al. (
2011) find that modifying analystbased
CofE metrics to account for “analyst sluggishness” improves the associations between some
CofE proxies and realized returns but does not always result in statistically reliable associations.
^{21} Other studies adopt a portfolio design: Gode and Mohanram (
2003), for example, find positive spreads in realized returns. In portfoliolevel tests, Hou et al. (
2012) show that returns spreads increase when they replace analyst forecasts with regressionbased earnings forecasts.
^{22} Li and Mohanram (
2014) use the Hou et al. (
2012) approach to derive earnings forecasts and show positive associations on the firm level as well.
3.2 Analysis of the CofE setting
We analyze a different and possibly coexisting explanation for the weak and inconsistent results in tests of association between realized returns and
CofE metrics. Rather than modify the metrics or consider other alternatives proposed in the literature, we leave the original
CofE metric constructions in place and propose an explanation that derives from known features of samples used to estimate those metrics. That is, we choose the most conservative starting point (the problem as first examined by Easton and Monahan (
2005)) and use unmodified implied
CofE definitions and unmodified realized returns.
Because the data requirements for estimating
CofE metrics eliminate firms with no analyst following, negative book value of equity, or negative or declining earnings forecasts, firms in a
CofE sample tend to be larger and more profitable, hence likely more stable, than the CRSP population. For example, Francis et al. (
2004, Table
1) report that the aggregate market capitalization of their sample of Value Linefollowed firms, averaging 790 firms per year, is just over 47% of the CRSP market capitalization. We posit these data requirements result in
CofE samples that are nonrandom draws from the population of CRSP firms, with a returns distribution that differs from the returns distribution in that population (or a random sample thereof).
^{23} We further posit that association estimates based on such a nonrandom sample are difficult to generalize to the population, that is, the external validity of the results is questionable. We do not dispute previous findings but rather use distributionmatching to arrive at results that we believe can be more justifiably generalized to the population of listed firms. To summarize, we believe the
CofErealized returns association has the following desirable characteristics for an empirical examination of the effects of nonrandom sampling: (1) the full distribution of returns is available for the reference sample; (2) data on the
CofE metric are missing for many reference sample firms, but conceptually all sample firms have a cost of equity; and (3) previous research shows the characteristics of returns for missing firms differ from the characteristics of returns for included firms.
To illustrate how data requirements may result in nonrandom samples, we let the requirements for five
CofE metrics dictate the sampling bias in returns. While intuition suggests these data requirements are likely to bias
CofE samples towards more stable firms with less dispersion in returns than the returns of the reference sample, intuition does not necessarily suggest an effect on associations of
CofE metrics with these returns. We triangulate the effects of nonrandom returns on associations by using implied risk factor premia that are estimable for the entire reference sample. Specifically, our data contain complete information on both realized returns and asset pricing factor betas (loadings) for all observations in the reference sample. In a CAPM world, crosssectional variation in beta is equivalent to crosssectional variation in expected returns. Therefore we can use association tests on factor loadings to gauge the nonrandomness of the
CofE sample by first performing factor loading association tests on the reference sample and then artificially imposing the
CofE sample restriction into the same test. Differences in results would suggest the
CofE sample is a nonrandom sample of the reference sample. Using FamaMacBeth twostage tests of the association of returns with risk factor betas, we show the
CofE returns sample is indeed a nonrandom sample of the reference sample.
4 Test design and nonrandomness of the CofE sample
4.1 Sample and descriptive data
Table
2 describes the archival data used in our empirical tests. We first identify all firms with monthly CRSP returns data during February 1976 to July 2009 (402 months). These data are used for our crosssectional asset pricing tests. The reference sample (the Full Returns sample) includes all firms with returns data in the current and preceding 11 months; a firm is required to have 12 consecutive monthly returns observations to enter the Full Returns sample in Month
t.
^{24} The Full Returns sample contains 2,460,998 firmmonth observations (24,657 unique firms). Table
2, Panel A, shows the average monthly cross section contains 6122 firms, with an average (median) monthly realized raw return of 1.30% (0.20%). Monthly excess returns, (realized return less the monthspecific riskfree rate) are 0.83% (mean) and −0.27% (median). The average crosssectional standard deviation of both raw and excess returns is 16.15%, and the interquartile ranges are about 12%–13%, indicating that realized returns are quite dispersed.
^{25}
Table 2
Descriptive statistics of monthly (crosssectional) distributions
# Firms

Mean

Std. Dev.

Skewness

Kurtosis

Min

P5

Q1

Median

Q3

P95

Max



Panel A: Full returns sample (asset pricing test sample)


Realized Returns

6122

0.0130

0.1615

3.7403

82.7487

−0.7442

−0.1947

−0.0595

0.0020

0.0676

0.2453

3.2195

Excess Returns

6122

0.0083

0.1615

3.7403

82.7487

−0.7488

−0.1994

−0.0642

−0.0027

0.0629

0.2406

3.2148

Panel B: Implied cost of equity sample


Realized Returns

955

0.0121

0.0896

0.6034

6.1594

−0.3734

−0.1217

−0.0393

0.0087

0.0594

0.1557

0.5742

Excess Returns

955

0.0074

0.0896

0.6034

6.1594

−0.3781

−0.1263

−0.0439

0.0041

0.0548

0.1510

0.5696

VL CofE

955

0.0121

0.0070

0.8016

2.7667

0.0002

0.0032

0.0061

0.0118

0.0165

0.0236

0.0513

GLS CofE

955

0.0071

0.0036

5.4237

59.8661

0.0006

0.0034

0.0053

0.0068

0.0084

0.0110

0.0505

MPEG CofE

955

0.0088

0.0045

3.2930

27.0187

0.0003

0.0037

0.0060

0.0081

0.0105

0.0160

0.0538

OJN CofE

955

0.0098

0.0037

4.5876

46.6087

0.0040

0.0061

0.0077

0.0092

0.0110

0.0153

0.0526

CT CofE

955

0.0073

0.0049

4.9134

49.7870

0.0001

0.0020

0.0046

0.0067

0.0090

0.0133

0.0580

Panel B of Table
2 reports average crosssectional statistics for the sample of firms with data to estimate the five
CofE measures. On average, those cross sections contain 955 firms (383,955 monthly observations for 3989 unique firms), a potentially nonrandom subsample of the Full Returns sample. Value Line cost of equity (
VL CofE) estimates are derived from Value Line target prices and dividend forecasts, are recalculated each month and are deannualized to the month level.
^{26} We calculate four other implied
CofE estimates following Claus and Thomas (
2001,
CT); Gebhardt et al. (
2001,
GLS); Easton (
2004,
MPEG), and Ohlson and JüttnerNauroth (
2005,
OJN).
^{27} The
CofE metrics require analyst following in general, and Value Line coverage in particular, as well as positive and increasing earnings forecasts.
As reported in Panel B of Table
2, the mean (median) values of the
CofE estimates range from 0.0071 to 0.0121 (0.0067 to 0.0118). The mean (median) monthly realized excess returns for the
CofE sample are 0.74% (0.41%). The Full Returns sample is more dispersed, more positively skewed, and more leptokurtic than the
CofE sample. With regard to dispersion, the standard deviation of excess returns for the
CofE sample is 8.96%, a 44.5% reduction compared to the standard deviation of the Full Returns sample, and the interquartile range of excess returns of the
CofE sample is 9.87%, a reduction of about 22% relative to the Full Returns sample. The Full Sample returns are positively skewed, with skewness coefficient of 3.74; the skewness coefficient of the
CofE sample is 0.6034 (a perfectly symmetric distribution has zero skewness). Finally, the Full Sample returns are leptokurtic, with a kurtosis coefficient of 82.75 on average, while the average
CofE sample kurtosis is 6.16.
4.2 Benchmark results of associations between CofE metrics and realized returns
We first estimate crosssectional Pearson correlations and slope coefficients from regressions of realized (excess) returns on each of the five
CofE metrics; intercepts (not tabulated, in the interest of brevity) are included in all regressions. We use Eq. (
6) to estimate slope coefficients for each Month
t using all complete returns
CofE observations available for that month.
$$ {R}_{i,t}{R}_{f,t}={\delta}_{0,t}+{\delta}_{1,t}{CofE}_{t1}+{\varepsilon}_{i,t}. $$
(6)
The averages of the monthly coefficient estimates
δ
_{1, t}over the sample period measure the association between realized excess returns and a specific
CofE metric. Following Fama and MacBeth (
1973), the test statistic for the significance of the associations is the average monthly coefficient estimate relative to the timeseries standard error of the monthly estimates over the sample period.
^{28} Results are reported in Table
3 for the unmodified
CofE sample resulting from deletion when data on any of the
CofE metrics is missing (analogous to Easton and Monahan
2005). These results are broadly consistent with prior literature on the association between
CofE estimates and realized returns, if not more negative.
^{29} Correlations show either no reliable relation between realized returns and
CofE, in the case of VL
CofE, or a negative relation, in the case of the other four
CofE metrics (tstatistics range from −0.52 to −6.11). Regression coefficients show a similar picture, with three of the five metrics showing significantly (at the .05 level or better) negative slope coefficients. All five coefficients are reliably different from their theoretical value of 1, with tstatistics (not tabulated) between −7.25 and −12.36.
Table 3
Average correlation and regression coefficients of realized returns and
CofE metrics
Actual
CofE Sample



Correlation

Regression


Coefficients

Coefficients


VL CofE

−0.0033

−0.0142

tstat

−0.52

−0.15

GLS CofE

−0.0096

−0.2273

tstat

−2.19

−1.34

MPEG CofE

−0.0203

−0.4121

tstat

−4.52

−3.61

OJN CofE

−0.0159

−0.3961

tstat

−3.44

−2.68

CT CofE

−0.0258

−0.4723

tstat

−6.11

−3.79

KS

0.1389


(
p value)

(0.0009)

4.3 Benchmark results of factor beta tests
To exploit the richness of the cost of capital setting we use tests of factor betas using twostage assetpricing tests; one purpose is to serve as an input to our demonstration that the
CofE sample is a nonrandom sample from the reference sample. In the first stage, we estimate slope coefficients (factor betas) in a firmspecific timeseries regression of excess returns on each risk factor:
where
R
_{i, t} −
R
_{f, t} is the excess return for firm
i for Month
t;
F
_{t} = a risk factor, specifically, the market excess return (market factor), the size factor or booktomarket factor (
SMB
_{t},
HML
_{t}) from FamaFrench (
1993) or the accruals quality factor (
AQfactor
_{t}) from Francis et al. (
2005);
\( {b}_{i,t}^F \) = the factor beta for risk factor
F;
t subscripts the sample month. Equation (
7a) is estimated over a rolling 12month window ending in Month
t.
^{30}
$$ {R}_{i,t}{R}_{f,t}={a}_{i,t}+{b}_{i,t}^F{F}_t+{\varepsilon}_{i,t}, $$
(7a)
In the second stage, we estimate crosssectional regressions of the firmspecific excess returns in Month
t on the univariate firststage factor loadings
\( \hat{b_{i,t}^F} \) (the risk factor betas) obtained from estimating Eq. (
7a):
$$ {R}_{i,t}{R}_{f,t}={\gamma}_{0,t}+{\gamma}_t^F\hat{b_{i,t}^F}+{\vartheta}_{i,t}. $$
(7b)
Equation (
7b) is estimated for each Month
t. The full sample tests use all firms with the necessary observations to estimate first stage betas. The secondstage coefficient estimates (
\( {\gamma}_t^F \)) are interpretable as implied risk factor premia in Month
t (implied by the firststage loadings). Following Fama and MacBeth (
1973), the test statistic for the significance of the implied risk factor premia is the average monthly coefficient estimate relative to the timeseries standard error of the monthly estimates over the sample period. Theory predicts the sign (positive), but not the magnitudes of the secondstage coefficient estimates (the magnitudes of the implied factor premia). Following previous research, we test whether the
\( {\gamma}_t^F \) estimates are reliably different from zero.
We use the samples described in Table
2 to establish benchmark associations between excess returns and factor betas. The tests are motivated by the idea that both
CofE metrics and factor betas are supposed to capture risk and the fact that tests on factor betas can be performed on
both the reference sample and on subsamples, such as the
CofE sample. Table
4, column 1, shows the second stage coefficient estimates from Eq. (
7b) and tstatistics based on the timeseries standard error of the monthly estimates. Our interest is not in the significance of specific risk factors but rather in using the Full Sample results as a benchmark for comparing subsample results. The association between returns and market beta is positive (the coefficient estimate corresponds to a risk premium of 0.52% per month; tstatistic = 2.03) as is the association for the
AQfactor beta (risk premium of 0.77% per month; tstatistic = 2.44). The coefficient on the
SMB beta is 0.0025, tstatistic = 1.69, significant at the 0.05 level, onetailed. The association between returns and
HML beta is not reliably different from zero at the 0.05 level.
^{31}
Table 4
Association tests on reference sample, random subsamples, and
CofE subsamples
Full returns sample

1000 Random subsamples (of monthspecific
CofE sample size)

Actual
CofE sample



Mean

Range


beta
^{Market}

0.0052

0.0050

[0.0032; 0.0069]

0.0022

tstat

2.03

1.93

[1.27; 2.56]

0.88

beta
^{SMB}

0.0028

0.0027

[0.0015; 0.0040]

0.0003

tstat

1.69

1.62

[0.93; 2.34]

0.20

beta
^{HML}

−0.0020

−0.0020

[−0.0030; −0.0010]

−0.0005

tstat

−1.33

−1.27

[−1.86; −0.63]

−0.32

beta
^{AQFactor}

0.0077

0.0074

[0.0051; 0.0090]

0.0023

tstat

2.44

2.32

[1.67; 2.79]

0.73

As previously described, we aim to shed light on how differences in the distributional properties of estimation samples of realized returns and, by implication, how differences in sample selection criteria affect the results of association tests between realized returns and both risk factor betas and
CofE estimates. We first consider sample size versus sample nonrandomness, using the Full Returns sample as the proxy for the population and the
CofE sample as a potentially nonrandom subsample. With regard to sample size, the monthly average is 6122 firms in the Full Returns sample and 955 firms in the
CofE sample, a reduction of about 84%. With regard to distributional properties, as captured by dispersion, skewness, and kurtosis, the Full Returns sample is more extreme on all three distributional properties.
Because factor betas are available for
both the Full Returns sample
and the
CofE sample, asset pricing tests can be used to illustrate that the
CofE sample differs from the Full Returns sample with respect to the association between realized returns and risk proxies. To illustrate the effects of sample size, columns 2 and 3 of Table
4 present results of association tests between risk factor betas and realized returns from 1000 randomly drawn subsamples of the Full Returns sample (Random Subsample) of the same size each month as the actual
CofE sample (monthly average of 955 firms).
^{32} Column 2 reports average slope coefficients and tstatistics, and column 3 contains the range of values across the 1000 random draws. The coefficient estimates of the Full Returns sample (column 1) and the average random subsample (column 2) are nearly identical (differences are between 1 and 4 basis points); the Full Returns results fall well into the range of values (column 3). The reduced sample size means the monthly coefficients are estimated with less precision; as expected, the timeseries tstatistics are lower by amounts between 0.10 (market risk premium) and 0.12 (
AQFactor premium). Turning to the effects of nonrandomness, results of the asset pricing tests using the actual
CofE sample are shown in the rightmost column of Table
4. None of the factor betas evidences a significant association with excess returns, and all factor premia are reduced in magnitude by at least 50%. The results from the
CofE sample fall outside the range of values spanned by the random subsamples.
We draw three conclusions from the results in Table
4. First, even substantial reductions in sample size (84% in the average cross section) have a modest effect on the efficiency of the estimation. Second, distributional differences in either realized returns or factor betas have substantial effects on the results. We interpret these results as supporting the notion that the
CofE sample is a nonrandom subsample of the Full Returns sample for purposes of testing associations with proxies for risk. Third, in such nonrandom samples, if factor betas fail to load significantly, insignificant results concerning
CofE metrics should not be surprising.
4.4 Demonstration of distributionmatching on simulated CofEcalibrated data and on archival data
4.4.1 CofEcalibrated simulations
Table
5 shows simulation results when the data approximate the size and shape of the distribution of the reference sample excess returns and the Value Line
CofE metric. We use empirically determined parameters of the actual distributions of excess returns and of a
CofE metric to create simulated data similar to the archival data. We are able to mimic the first four moments of the variable distributions. We induce correlations of 0.25, 0.10, and 0. We report results for NonRandom Sample I, constructed to be less extreme than the random sample by setting the selection probability to decrease in the absolute distance to the variable mean. For the simulated Full Returns data (first line of Table
5) and for random samples of the same size as the actual
CofE sample (second line), estimated correlations are similar to the induced population correlations. This finding buttresses our conclusion that even sharply diminished sample sizes do not obscure or shift estimates away from true correlations in the data, as long as the samples are randomly drawn from the reference sample. In contrast, the third line indicates that for the intentionally lessextreme nonrandom sample, the KS test statistic rejects similarity of the distributions at better than the 0.0001 level. Estimated associations between the simulated returns and simulated
CofE metrics are negative and highly significant even though true correlations are positive: when the true correlation is 0.25 (0.10), the estimated correlation is −0.17 (−0.06), with a tstatistic of −61.8 (−33.9). When the true correlation is zero, the estimated correlation for the nonrandom sample is also zero (point estimate 0.0004, tstatistic 0.23). When we distributionmatch the nonrandom sample, the KS test statistics decline to about 0.03 with significance levels of about 0.50. When the true correlation is 0.25 (0.10) [0], the estimated correlation is 0.17 (0.07) [0.00], with a tstatistic of 25.40 (9.07) [0.30].
Table 5
Simulation results from
CofEcalibrated samples
CORR* = 0.25

CORR* = 0.10

CORR* = 0



KS

CORR

KS

CORR

KS

CORR


Full “Returns” Samples

N/A

0.2503

N/A

0.1000

N/A

0.0001

Random Samples (m = CofE sample size)

0.0262

0.2503

0.0265

0.1005

0.0263

0.0005

(p value) 
tstat

(0.6580)

134.98

(0.6504)

58.46

(0.6590)

0.31

NonRandom Sample (m = CofE sample size)


Ranked abs. distance to mean

0.1847

−0.1662

0.1844

−0.0627

0.1841

0.0004

(p value) 
tstat

(0.0000)

−61.80

(0.0000)

−33.91

(0.0000)

0.23

DistributionMatched Sample

0.0319

0.1737

0.0319

0.0659

0.0318

0.0023

(p value) 
tstat

(0.5017)

25.40

(0.4974)

9.07

(0.5015)

0.30

We believe these simulations support two conclusions. First, sample marginal distributions, not sample size per se, affect the ability to empirically detect the true correlations between two variables. In particular, the sign differences reported in Table
5—negative correlation estimates when true correlations are positive—highlight the potentially substantial bias in results when the marginal distribution is nonrandomly drawn from the reference sample. Second, despite extreme differences in the characteristics of marginal distributions, distributionmatching yields an adjusted sample with correlations similar in sign and magnitude to the true correlations.
4.4.2 Distributionmatching the actual CofE sample
We construct distributionmatched samples from data on realized returns and the five
CofE metrics. Results so far suggest the returns of the actual
CofE sample are a nonrandom sample of the Full Returns sample returns. Table
2 shows excess returns for the
CofE sample have a similar mean/median, and noticeably smaller standard deviation, skewness and kurtosis, as compared to the Full Returns sample. We interpret these findings as raising the question of whether the negative or weak correlation between
CofE estimates and realized returns reported in Table
3 is generalizable to the reference sample or arises from the effects of data requirements.
To accommodate the substantial differences in the shape of the
CofE sample returns distribution, as compared to the reference sample returns distribution, we change the implementation of the distributionmatching approach used in the simulation in two ways. We first apply an iterative procedure that starts with a base sample, draws an additional observation, and keeps that additional observation only if the resulting sample shows a smaller KS statistic. The approach still aims to minimize the KSbased statistic, even in months where insignificant KS statistics cannot be achieved because of large initial differences between the
CofE sample and Full Returns sample.
^{33} The minimization does not require a prespecified sample size but rather lets the iteration determine the optimal sample when the KS statistic cannot be further minimized. This approach is conceptually grounded but inefficient and computationally burdensome for large samples. The second, less computationally demanding implementation matches the
CofE sample to the Full Returns sample using a variant of stratified resampling (poststratification), which tries to match the dispersion of the returns distribution. We describe both approaches next.
Method 1: KolmogorovSmirnovbased distributionmatch
We start by randomly sampling either 20% of the monthspecific sample or, separately, 100 unique firms from the
CofE sample in a given month. We compute the KS statistic for this initial draw against all returns from the reference sample in that month.
^{34} Our previous analyses suggest the KS test on this initial sample is likely to reject the hypothesis that the Full Returns distribution is equal to the returns distribution of, for example, the 100 initially selected firms. We start our iteration to minimize the KS statistic by randomly adding one observation (# 101), recompute the KS statistic, and again compare to the reference distribution of returns that month. If the KS statistic using 101 observations (against the reference distribution) is equal to or greater than the KS statistic using the original 100 observations sample (against the reference distribution), we dismiss the 101st observation and replace it with another randomly chosen, with replacement, 101st observation from the
CofE sample. If the KS statistic using 101 observations is lower than the KS statistic using the 100 observations sample, we keep the 101st observation, draw a 102nd observation, and evaluate the inclusion of the 102nd observation. We repeat this step 1000 times, thereby allowing for KSbased distributionmatched samples to increase by a maximum of 1000 observations each month.
^{35} Because convergence to a minimum KS statistic depends both on the initial 20% (or 100) observations drawn and on the order of additions, we repeat the procedure 30 times and retain the final sample with the lowest KS statistic (i.e., with the minimal difference as compared to the Full Returns distribution). We repeat the construction of KSbased samples for each month. When the iteration begins with 20% of the
CofE sample firms, the final distributionmatched sample contains an average of 242.2 firms (about 25.4% of the 955 firms in the average
CofE sample month), with a timeseries standard deviation of 54 firms. When the iteration begins with 100 firms each month, the average cross section consists of 124 firms (with a standard deviation of 14 firms).
^{36}
Method 2: Description of binbased distributionmatch
To reduce the computational burden of Method 1, we create binbased distributionmatched samples. Binbased matching is similar to poststratification except that the distribution is also truncated (some population strata are not represented in the sample), requiring an additional weighting scheme for the tails of the distribution. For the common support region, we first place the returns of the Full Returns sample and the
CofE sample into bins with width of 100 basis points (bp) and calculate the sample proportion of observations in each bin for both samples.
^{37} To distributionmatch, we reweight (by resampling return
CofE pairs with replacement) each bin in the
CofE sample, so that the sample proportion matches the proportion in the corresponding reference sample bin. For example, if the realized returns bin [0.10; 0.11[contains 5% of the
CofE sample observations and 10% of the reference sample observations, we resample the
CofE sample bin to increase its percentage to 10% of the sample size in that month.
^{38} At the extremes of the reference sample distribution, we encounter bins without corresponding observations in the
CofE sample. To address this issue, at both the upper and lower extremes of the
CofE sample, we reweight the most extreme positive and most extreme negative returns, with equal weighting at both extremes in the following form (month subscripts omitted):
$$ {w}_{i_{CofE}}=\Big\{{\displaystyle \begin{array}{c}{w}_{i_{RS}}\sum \limits_{j=\min \left({i}_{RS}\right)}^{\min \left({i}_{CofE}\right)}{\left[\mathit{\min}\left({i}_{CofE}\right){i}_{j, RS}+1\right]}^{\gamma}\kern1.08em \forall \kern0.48em {i}_{RS}=\min \left({i}_{CofE}\right)\\ {}{w}_{i_{RS}}\kern8.399994em \forall \kern0.48em \min \left({i}_{CofE}\right)<{i}_{RS}<\max \left({i}_{CofE}\right)\\ {}{w}_{i_{RS}}\sum \limits_{j=\max \left({i}_{CofE}\right)}^{\max \left({i}_{RS}\right)}{\left[{i}_{j, RS}\max \left({i}_{CofE}\right)+1\right]}^{\gamma}\kern0.72em \forall \kern0.48em {i}_{RS}=\max \left({i}_{CofE}\right)\end{array}} $$
(8)
\( {w}_{i_{CofE}} \)(
\( {w}_{i_{RS}} \)) is the sampling proportion of Bin [
i;i+0.01] in the
CofE sample and the reference sample, respectively. The product of sampling proportion
\( {w}_{i_{CofE}} \) and overall sample size in Month
t is the binspecific number of draws that month. We numerically solve, within the sampling procedure, for the constant weight parameter
γ until the standard deviation for the distributionmatched
CofE sample is statistically indistinguishable from that of the Full Returns sample. After this calibration, the timeseries average of the differences in crosssectional standard deviations between the Full Returns sample and the distributionmatched
CofE sample is 0.0009 (tstatistic = 0.62) at γ = 2.15, and −0.0008 (tstatistic = −0.56) at γ = 2.20. Figure
2a illustrates the approach by plotting the average distribution of excess returns for the
CofE sample before and after distributionmatching as well as the reference distribution of returns. The dashed continuous distribution of returns in the
CofE sample is sorted into bins of predetermined widths and then resampled, such that the sample proportion of each bin is equal to that bin in the reference distribution. The procedure is effective if the heights of the resulting light bars, representing the strata, match the heights of the dark reference strata. As previously mentioned (note 10), Fig. 2b plots the average Value Linebased
CofE metric by returnsbin, before and after binbased distribution matching; across the 402 sample months, in none of the 101 returns bins is the difference significantly different from zero at the 0.01level (results not tabulated). Differences using the other
CofE metrics (not plotted) are, if anything, generally smaller in absolute magnitude.
×
Association tests using distributionmatched samples
Table
6 reports results of correlation tests between realized returns and
CofE metrics for the distributionmatched samples under both the KolmogorovSmirnov and binbased approaches. The average KS statistic (average
p value) of the unadjusted
CofE sample is about 14% (0.0009) and the test rejects similarity of distributions in 401 of 402 sample months at the 0.10 level or lower. After KSbased distributionmatching using 20% of firms (100 firms), the average KS statistic is just under 6% for both initializations, with an average
p value of 0.5287 (0.7789), and the test rejects similarity of distributions in 42 (5) of 402 months at the 0.10 level or better. We conclude from these results that the KSbased approach to distributionmatching is effective.
Table 6
Association tests in distributionmatched samples
DistributionMatched
CofE Samples



KSBased Sampling

Binbased Weighted Sampling


Initial # = 20%

Initial # = 100

γ = 2.15

γ = 2.20


Avg. KS Statistic

0.1389

0.0581

0.0589

0.0992

0.1031

Avg. p value

0.0009

0.5287

0.7789

0.0280

0.0248

# Months with p ≤
0.10

401

42

5

372

375

Average crosssectional correlation coefficients


VL CofE

−0.0033

0.0524

0.0496

0.0514

0.0537

tstat

−0.52

6.08

5.35

4.14

4.28

GLS CofE

−0.0096

0.0312

0.0278

0.0399

0.0413

tstat

−2.19

4.53

3.52

3.85

3.95

MPEG CofE

−0.0203

0.0270

0.0274

0.0286

0.0305

tstat

−4.52

3.98

3.59

2.54

2.67

OJN CofE

−0.0159

0.0316

0.0289

0.0350

0.0370

tstat

−3.44

4.36

3.57

3.10

3.23

CT CofE

−0.0258

0.0153

0.0057

0.0213

0.0227

tstat

−6.11

2.23

0.73

2.07

2.18

Average crosssectional regression coefficients


VL CofE

−0.0142

0.9000

0.9160

1.0828

1.1394

tstat (against 0)

−0.15

5.46

5.07

3.42

3.51

tstat (against 1)

−10.68

−0.61

−0.46

0.26

0.43

GLS CofE

−0.2273

1.3155

1.0740

2.2834

2.3356

tstat (against 0)

−1.34

3.93

2.70

3.30

3.30

tstat (against 1)

−7.25

0.94

0.19

1.85

1.89

MPEG CofE

−0.4121

0.7904

0.7075

1.0727

1.1192

tstat (against 0)

−3.61

3.62

2.83

2.10

2.13

tstat (against 1)

−12.36

−0.96

−1.17

0.14

0.23

OJN CofE

−0.3961

1.1476

0.8862

1.6353

1.6930

tstat (against 0)

−2.68

3.98

2.63

2.59

2.62

tstat (against 1)

−9.45

0.51

−0.34

1.01

1.07

CT CofE

−0.4723

0.5082

−0.0143

1.0092

1.0418

tstat (against 0)

−3.79

1.99

−0.05

2.00

2.02

tstat (against 1)

−11.83

−1.93

−3.45

0.02

0.08

The top portion of Table
6 reports correlations between five
CofE measures and realized returns, using the KSbased distributionmatched samples (columns 2 and 3) and the binbased distributionmatched samples (columns 4 and 5). For KSbased matching, the timeseries average correlations across 402 months are reliably positive in 9 of the 10 specifications, with tstatistics between 2.23 and 6.08 (the exception is the association with the CT
CofE metric with 100 firms as initial sample where the correlation is positive and insignificant). These results indicate reliably positive associations between four implied
CofE metrics and realized returns, in contrast to the benchmark results on the unmodified
CofE sample (reproduced in the first column), where four correlations are significantly negative. For binbased matching, we report correlations for γ = 2.15 and γ = 2.20, where the overall difference in standard deviations is insignificant at conventional levels. Because the focus is on matching standard deviations only, we do not expect the binbased distributionmatch to be entirely effective in lowering the KS statistics for general similarity of distributions. Table
6, columns 4 and 5, shows that the average KS statistic decreases, relative to the unmodified
CofE sample, and remains significant for 372 (375) months for γ = 2.15 (γ = 2.20). In these analyses, all five
CofE measures have reliably positive correlations with realized returns, ranging from 0.021 (CT
CofE measure) to 0.054 (the VL measure), with tstatistics between 2.07 (CT
CofE measure, γ = 2.15) and 4.28 (VL
CofE measure, γ=2.20).
The bottom portion of Table
6 contains regression coefficients for the KSbased and binbased distributionmatched samples. In contrast to the results reported in Table
3, the coefficients are generally significantly positive, with tstatistics usually exceeding 2.0, and statistically indistinguishable from 1 in 16 of the 20 specifications. The coefficient on the
CT metric is significantly smaller than 1 in two of the KSbased samples, and the
GLS CofE metric shows significantly larger coefficients for the binbased samples.
To examine the effect of distributionmatching on the implied factor premia from asset pricing tests, we repeat the FamaMacBethtype tests reported in Table
4 using the KSbased distributionmatched samples with the initialization set at 20% of the
CofE sample (results not tabulated) and using one binbased distributionmatched sample (γ = 2.20; results not tabulated). The factor premia from these samples are qualitatively similar to the Full Returns sample results in Table
4 in that the market factor,
SMB factor, and
AQfactor are positive and statistically significant and the
HML factor is insignificant at conventional levels. In sum, the results on implied factor premia in the distributionmatched samples are qualitatively similar to results in the reference sample as reported in Table
4.
We next address a concern that arises in part from Botosan et al.’s (
2011) finding that newspurged realized returns, which should measure expected returns, have either no associations or counterintuitive associations with risk proxies such as beta. In our setting, the concern is that distributionmatching the
CofE sample increases the association between
CofE metrics and excess returns at the cost of a diminished association between
CofE metrics and other risk proxies, specifically risk factor betas. We test for a decline in the associations between
CofE metrics and risk factor betas, using (1) the sample composition from the KSbased matching in Table
6 with initial sample size equal to 20% of the
CofE sample, as compared to (2) a random sample from the
CofE sample of the same size in any given month. For both samples, we regress the five
CofE metrics on lagged risk factor betas from Eq. (
7a). If distributionmatching decreases the association between
CofE metrics and risk factor betas, the associations will be smaller for the distributionmatched sample (1) than for the random sample (2). Our test is based on the timeseries of the difference between the 402 monthspecific KSbased sample results and the 402 monthspecific results from the equalsized random samples. We repeat the procedure 100 times and evaluate the differences using the average FamaMacBethtype tstatistics across the 100 outcomes. In untabulated results, we find that for 19 of 20 coefficient estimates (five
CofE metrics times four risk factor betas), differences between the two sets of associations are insignificant at conventional levels, with tstatistics between −0.59 and 1.46. The exception is the coefficient on the market beta in the VL
CofE regression, which shows a small, reliably positive difference of 0.0001 (
t = 2.09). In all cases, coefficients from the KSbased sample are numerically similar to coefficients from the random samples; they are always of the same sign and always significant at comparable levels.
Combined with previous results, we interpret the weight of the evidence in Table
6 as demonstrating that differences in the shape of the returns distribution between the Full Sample and the
CofE sample have a marked effect on the results of association tests. We draw three inferences from these results. First, selection criteria that yield estimation samples with different returns distributions, as compared to a reference sample, decrease the ability to detect theoreticallypredicted associations between realized returns and
CofE estimates. Second, adjusting the distribution of the outcome variable (in this case, realized returns) in the nonrandom sample to mimic that of the reference sample provides at least a partial solution. Third, the finding that the distributionmatched sample may be smaller than the original, nonrandom sample suggests that attempts to achieve generalizability to a reference sample by maximizing the size of a datarestricted sample may not be effective.
5 Extensions
5.1 Relating distributionmatching to selection models and multiple imputation
Selection models
Both selection models and distributionmatching seek to incorporate information into the test model beyond the information in the completedata subsample. The latter focuses only on the outcome variable, whose empirical distribution in the reference sample can be derived by the researcher or is empirically estimable, and aims to construct a test sample that appears randomly selected with respect to the outcome variable. In contrast, a selection model (1) operates under the assumption that data are
not missing at random, conditional on observed data, and (2) requires explicit modelling of the missingness mechanism using additional explanatory variables, which might impose even more stringent data restrictions than the actual test model. Thus the sample restriction issue at the heart of our analysis does not arise in the approach developed by Heckman (
1979), because the exogenous covariates in the firststage selection model are attainable for all observations, or, equivalently, attainable for a
random subsample of the population.
^{39} Consequently, results from a Heckmantype model are generalizable only to the sample for which the selection model variables are available, and increasing selectionmodel fit by including more explanatory variables is likely to impose increasingly stringent sample restrictions due to data requirements. Exacerbating the data availability problem is the exclusion restriction on the explanatory variables set in the test model, compared to the explanatory variables set in the selection model. To avoid collinearity of the test model variables and the inverse Mills ratio, the recommended approach is to include at least one additional variable in the selection model not contained in the test model of interest and, in theory, not associated with the outcome variable.
^{40}
Distributionmatching is, by design, nonparametric and based on a reference distribution of the outcome variable. In contrast, the derivation of the Heckman correction for sampling biases relies on the assumption that the residuals from the selection model and the test model are jointly normally distributed. The normality assumption allows for a closedform solution for the sampling bias in OLS coefficients as a function of the inverse Mills ratio, the standard deviation of the test model residual, and the correlation between test model residuals and selection model residuals. The normality assumption is crucial for the parameter estimates in the test model; descriptive statistics for excess returns reported in Table
2 cast doubt on this assumption in our setting. At a minimum, we caution that Heckman test results with realized returns as the dependent variable are likely biased (in an unknown direction) by violations of the normality assumption.
Despite these concerns, we implement the Heckman model, subject to the constraint of avoiding, as much as possible, additional sample restrictions, at the potential cost of not maximizing the selection model fit. We restrict our analysis to selection models with explanatory variables available for all, or at least the vast majority of, observations in the Full Returns sample and include some or all of the following: firm size (CRSP market capitalization at the end of the prior month), firm age (number of months between the first month on CRSP and the current month), CRSP trading volume, the booktomarket ratio (calculated from the Compustat annual file), and the four univariate risk factor betas from the asset pricing regressions, eq. (
7a).
^{41}
Table
7 reports semipartial correlation coefficients between
CofE metrics and excess returns and goodnessoffit measures for the probit selection models estimated. The model including only size has a pseudoR
^{2} of 0.48, with no additional sample loss; adding variables increases the pseudoR
^{2} to a maximum of 0.52, with a sample loss of 2.1% when the model includes the log of the booktomarket ratio. The reported semipartial correlations are averages of the 402 monthspecific (crosssectional) semipartial correlations, obtained by controlling for the inverse Mills ratio in the respective
CofE metric first and then computing the correlation between the returns and the residualized
CofE metric. Similarly, regression coefficients (bottom portion of the table) are averages of 402 crosssectional slope coefficients from regressions of excess returns on both the
CofE metric in question and the inverse Mills ratio from the selection model.
Table 7
Results from Heckmantype selection models
No correction

Lag (MktCap)

Lag (MktCap), Volume, Age

Lag (MktCap), Volume, Age, B/M

Factor Betas

Combined



(I)

(II)

(III)

(IV)

(V)

(VI) = (IV) + (V)


Average crosssectional semipartial correlation coefficients


VL CofE

−0.0033

−0.0079

−0.0060

−0.0067

−0.0371

−0.0055

tstat

−0.52

−1.32

−0.98

−1.10

−8.79

−0.91

GLS CofE

−0.0096

−0.0146

−0.0118

−0.0126

−0.0254

−0.0117

tstat

−2.19

−3.57

−2.81

−3.02

−8.34

−2.81

MPEG CofE

−0.0203

−0.0266

−0.0235

−0.0244

−0.0335

−0.0234

tstat

−4.52

−6.49

−5.59

−5.81

−11.25

−5.70

OJN CofE

−0.0159

−0.0215

−0.0190

−0.0200

−0.0309

−0.0191

tstat

−3.44

−5.00

−4.32

−4.55

−10.15

−4.42

CT CofE

−0.0258

−0.0299

−0.0279

−0.0286

−0.0342

−0.0278

tstat

−6.11

−7.73

−6.98

−7.24

−11.98

−7.15

Average crosssectional regression coefficients


VL CofE

−0.0142

−0.0760

−0.0583

−0.0667

−0.3470

−0.0806

tstat (against 0)

−0.15

−0.84

−0.63

−0.73

−5.23

−0.93

tstat (against 1)

−10.68

−11.85

−11.51

−11.67

−20.29

−12.46

GLS CofE

−0.2273

−0.3867

−0.3321

−0.3478

−0.6980

−0.3807

tstat (against 0)

−1.34

−2.35

−2.03

−2.15

−5.63

−2.47

tstat (against 1)

−7.25

−8.43

−8.15

−8.33

−13.69

−8.96

MPEG CofE

−0.4121

−0.5387

−0.5011

−0.5195

−0.7455

−0.5494

tstat (against 0)

−3.61

−5.11

−4.67

−4.87

−9.56

−5.62

tstat (against 1)

−12.36

−14.59

−13.99

−14.24

−22.39

−15.84

OJN CofE

−0.3961

−0.5362

−0.4991

−0.5213

−0.8506

−0.5527

tstat (against 0)

−2.68

−3.84

−3.54

−3.71

−8.05

−4.25

tstat (against 1)

−9.45

−11.01

−10.62

−10.84

−17.52

−11.93

CT CofE

−0.4723

−0.5529

−0.5308

−0.5464

−0.7797

−0.5642

tstat (against 0)

−3.79

−4.73

−4.48

−4.65

−8.48

−5.13

tstat (against 1)

−11.83

−13.28

−12.92

−13.17

−19.35

−14.23

Auxiliary Information


Avg. Pseudo R
^{2}

N/A

0.48

0.51

0.51

0.05

0.52

Avg. Sample N

955

955

943

939

955

939

Avg. Sample Loss (%)

N/A

0.0%

1.7%

2.1%

N/A

2.1%

Avg. Reference Sample N

6122

6117

5670

4883

6122

4883

Avg. Reference Sample Loss (%)

N/A

0.1%

10.0%

22.4%

N/A

22.4%

The inverseMills ratioadjusted semipartial correlations and the adjusted regression coefficients are negative or, in the case of the VL
CofE metric, indistinguishable from zero. Across
CofE metrics, point estimates appear slightly lower and are more statistically different from zero, as compared to unadjusted correlations or slopes.
^{42} With the caveat that the effects of the inverse Mills ratio on the semipartial correlations and regression coefficients might be due to violations of the normality assumption, an inadequate fit of the selection model, or some combination of the two, we conclude that Heckmantype selection models do not change the conclusion from results obtained using the unadjusted
CofE sample.
Multiple imputations
^{43}A standard implementation of multiple imputation will fail to recognize differences in the functional form connecting returns and
CofE. Specifically, correlation and regression coefficients from a
single imputation model for the entire crosssection of returns are qualitatively similar to the actual
CofE sample results reported in Table
3, except that coefficients tend to be more negative (farther from the theoretical value) and standard errors are larger because of additional variance from the imputed
CofE data. However, when we modify the approach to allow for groupwise imputation models (two groups divided at the monthly crosssectional median; three groups or five groups) to improve the overall model fit, 12 of the 15 regression slopes (five
CofE metrics times 3 different sample groupings) are positive, significant at the 0.10 level or better and indistinguishable from 1. Results for correlations are generally positive but weaker and insignificant in eight of the 15 specifications and especially for the
CT CofE.
We assess the sensitivity of these results in two ways. First, we preclude the imputation of negative values
^{44} by using log transformations before imputing and find qualitatively comparable results for two and three imputation groups and stronger results for five imputation groups per cross section. Qualitative inference changes only for the MPEG
CofE, with a higher coefficient of 0.65 (tstatistic against 0 = 1.83, tstatistic against 1 = −0.97). Second, we impute data for the factor tests. We construct a dataset that deletes the loadings estimates from eq. (
7a) for observations without
CofE metrics and then impute the now missing (by construction) values for the loadings before we estimate the implied factor premia using eq. (
7b). Untabulated results show that, for imputations of the full cross section, only the market risk premium is significantly positive (t=1.92). For imputations using two, three and five groups results are qualitatively similar to the results from the Full Returns sample. We interpret the weight of this evidence as suggesting that multiple imputation can be a viable alternative to distributionmatching, albeit one that imposes a normality assumption and that may require additional adjustments in a specific research setting, for example, precluding inadmissible imputed values.
5.2 Asset pricing tests on returns of samples that meet selection criteria used in accounting research
Using the CRSP population of firms with at least 12 consecutive monthly returns during our sample period and the subsample of those returns associated with firms for which
CofE measures can be calculated, we have analyzed how differences in returns distributions between the two samples affect results of association tests. We next consider whether results of asset pricing tests of the association between risk factor betas and realized returns are sensitive to the following crosssectional selection criteria that likely yield nonrandom samples: S&P 500 membership, a potential screen in compensation research
^{45}; NYSE listing, a screen in some intraday trading studies; availability of the standard deviation of analysts’ earnings forecasts, required for research examining forecast dispersion; or a stock price of at least $5. We apply each criterion separately to the Full Sample, report the proportions of firms that do and do not meet the criterion and reestimate Eq. (
7b), separately, for observations meeting and not meeting the criterion.
Results are reported in Table
8, Panel A. The selection criteria generally result in unequal proportions of firms in the Full Returns sample that do and do not meet each criterion. The difference in proportions is, unsurprisingly, most extreme for the S&P 500 criterion (8.44% meet the criterion). KS statistics for tests of equality of distributions show that, for three of the four selection criteria, percentage deviations between the Full Returns sample and the subsample meeting the criterion exceed the deviations for the subsample not meeting the criterion. That is, the returns distributions of firms
not selected by these three criteria more closely resemble the returns distribution of the Full Returns sample.
^{46} The exception is the price of at least $5 criterion.
Table 8
Univariate associations between (excess) returns and risk factor betas (implied factor premia)
Panel A: CrossSectional Sample Selection Criteria


S&P500 Member

Listed on NYSE

σ(EPS forecasts) missing

Price at least $5


No

Yes

No

Yes

No

Yes

No

Yes


beta
^{Market}

0.0053

0.0017

0.0059

0.0029

0.0023

0.0061

0.0061

0.0042


2.08

0.65

2.27

1.14

0.96

2.33

2.28

1.71


beta
^{SMB}

0.0029

−0.0007

0.0031

0.0008

0.0011

0.0034

0.0031

0.0029


1.74

−0.43

1.86

0.45

0.70

2.02

1.89

1.74


beta
^{HML}

−0.0021

−0.0001

−0.0023

−0.0007

−0.0020

−0.0023

−0.0023

−0.0018


−1.38

−0.04

−1.48

−0.45

−1.31

−1.50

−1.46

−1.21


beta
^{AQFactor}

0.0078

0.0010

0.0083

0.0030

0.0035

0.0089

0.0099

0.0079


2.47

0.30

2.62

0.93

1.13

2.80

3.05

2.57


Avg. Proportion of Firms

91.56%

8.44%

67.56%

32.44%

41.34%

58.66%

24.40%

75.60%


Avg. Number of Firms

5621

500

4131

1991

2641

3520

1497

4625


Avg. KS statistic

0.0129

0.1413

0.0454

0.0955

0.0761

0.0521

0.1655

0.0540


Avg. p value

(0.7109)

(0.0051)

(0.0264)

(0.0022)

(0.0055)

(0.0471)

(0.0000)

(0.0232)


Panel B: Induced Variation in the (Marginal) Distribution of Excess Returns


Induced Sampling Change in Positive Excess Returns (Right Tail)


−25%

−10%

−5%

−2.50%

0%

+2.50%

+5%

+10%

+25%


Avg. Monthly Proportion

23.72%

38.21%

43.17%

45.66%

48.15%

50.65%

53.15%

58.15%

73.07%

beta
^{Market}

0.0028

0.0045

0.0047

0.0050

0.0052

0.0053

0.0054

0.0059

0.0072

1.36

1.92

1.93

1.98

2.03

2.02

2.03

2.13

2.50


beta
^{SMB}

0.0005

0.0020

0.0024

0.0026

0.0028

0.0030

0.0033

0.0037

0.0054

0.38

1.29

1.48

1.59

1.67

1.77

1.93

2.13

2.98


beta
^{HML}

−0.0008

−0.0017

−0.0018

−0.0019

−0.0020

−0.0022

−0.0022

−0.0024

−0.0031

−0.66

−1.16

−1.19

−1.27

−1.32

−1.37

−1.40

−1.50

−1.83


beta
^{AQFactor}

0.0018

0.0052

0.0064

0.0071

0.0076

0.0082

0.0089

0.0103

0.0148

0.66

1.74

2.07

2.28

2.42

2.56

2.75

3.12

4.42

These findings suggest asset pricing tests may yield results that are more theoryconsistent as well as more consistent with results for the Full Sample for firms
not included in the sample resulting from the application of plausible selection criteria. Specifically, Table
8 shows that both point estimates and tstatistics are more similar to Full Sample results for firms that do
not meet the selection criteria. We view these results as indicative, but not dispositive, that the distributional issues we identified and analyzed for the
CofE sample generalize to other research situations where dataconstrained samples consist of large, stable firms and, as a consequence, have returns distributions that are not random draws from the population.
5.3 Association tests between realized returns and factor betas using forced nonrandom samples
To illustrate the sensitivity of association test results to (small) changes in nonrandom sampling, we split the Full Sample realized returns distribution into positive and negative returns and reweight both subsamples differentially to varying degrees. This test is motivated by the conjecture that data requirements might lead (implicitly) to a similar and less extreme reweighting of positive and negative returns. We resample 20 times, by month, for each of 402 sample months. In each month, with
N
_{t} firms in each month, we resample 20 times with replacement
N
_{t} firms. We use these 402 months of resampled data to illustrate the effect on the returnsfactors betas association, as these data are available for the entire reference sample.
Results are reported in Table
8, Panel B. The 0% column shows results when we resample preserving the population proportions of positive and negative returns. These results coincide with the Table
4 Full Returns sample results; small differences result from sampling with replacement as opposed to using the full sample. The columns labeled −2.5%, −5%, −10%, and −25% show the Full Returns sample results when our resampling procedure decreases the portion of positive returns sampled by the specified percentages and increases the portion of negative returns sampled by the same percentages. The columns labeled +2.5%, +5%, +10%, and +25% show results when resampling increases (decreases) the portion of positive (negative) returns sampled by the specified percentages. The results suggest that increasing the proportion of positive returns increases the significance of results of asset pricing tests.
^{47} For example, the tstatistic on the implied
SMB factor premium increases from 0.38 (25% decrease in positive returns) to 1.67 (unbiased sample) to 2.98 (25% increase in the proportion of positive returns). Factor premia are differentially sensitive to these changes, with the market factor apparently relatively more robust compared to other factors, although the trend exists also for it. We infer that results of association tests are sensitive to the distributional properties of estimation samples and therefore sensitive to differences in sample selection criteria, with the degree of sensitivity differing with the nature of the selection criteria.
Overall, the results in Table
8, Panel B, illustrate that the outcome of all four returnsbetas association tests is sensitive to a prespecified characteristic of the returns distribution, namely whether the sample contains more positive returns. We directly manipulated the sample distribution; however, this sample characteristic may also be implicitly influenced by researcherchosen selection criteria, data availability, or sample partitions.
5.4 Applying distributionmatching to Richardson et al. (2005)
We apply distributionmatching to Richardson et al.’s (
2005) analysis of the association between
annual returns and accruals to illustrate empirically that distributionmatching does not produce false results.
^{48} Richardson et al. find results consistent with prior research (notably Sloan
1996) and with their theory that lowerreliability accruals lead to lower earnings persistence that investors appear to not fully understand. We therefore expect that applying distributionmatching should similarly produce results consistent with prior research and theory; that is, distributionmatching should not falsify or bias these results. We follow procedures outlined in Section
3 and footnote 8 of Richardson et al. to obtain a sample as close as practicable to theirs
^{49} (the unadjusted accruals sample) and replicate the analysis presented in their Table
8, Panel B. Before distributionmatching, the average crosssectional KS statistic rejects similarity of the $5pricefiltered reference distribution of returns (as described in footnote 49) and the accrualssample distribution of returns at the 0.0917 level; differences are significant in 30 of the 40 sample years. After applying the KSbased distributionmatching approach, the average annual KS statistic is reduced to 0.0525, with an average
p value of .6377. Thus the sample of returns obtained by requiring specified accounting data appears to be nonrandom relative to the $5pricefiltered reference distribution of returns, and distributionmatching lets the distribution approximate a randomlydrawn sample distribution. The distributionmatched sample has fewer observations (approximately 23,000 as opposed to over 105,000 in the unadjusted accruals sample).
The key test variables reported in Richardson et al.’s Table
8, Panel B, are ROA (coefficient = 0.09, t = 1.69), change in working capital (∆WC; coefficient = −0.30, t = −7.54), change in net noncurrent operating assets (∆NCO; coefficient = −0.27, t = −6.77) and change in net financial assets (∆FIN; coefficient = −0.054, t = −1.94). After distribution matching, we find the following (not tabulated): the coefficient on ROA is 0.16, t = 2.27; the coefficient on ∆WC is −0.31, t = −4.28; the coefficient on ∆NCO is −0.17, t = −3.01; the coefficient on ∆FIN is −0.02, t = −0.29. Using twosample ttests of differences in slope coefficient estimates from the full sample and the distributionmatched sample, we do not reject equality at lower than the 0.18 level. In other words, the distributionmatched sample yields results that are statistically indistinguishable from the fullsample results of Richardson et al.’s returns test. We believe these results complement our previous analyses by supporting an inference that distribution matching does not produce false results.
5.5 Eliminating the requirement of Value Line data
We reestimate the Table
6 association tests after dropping the requirement that
CofE firms be followed by Value Line (results not tabulated). For these tests, the
CofE sample contains firms with the necessary IBES data to calculate four
CofE metrics; the monthly average is 1980 firms, a substantial increase from the monthly average of 955 firms when we impose the Value Line requirement.
^{50} However, the larger IBES sample returns distribution remains reliably different from the reference sample of returns: the KS statistic for the unadjusted IBES
CofE sample is 11.21% with average p value = 0.0018; 400 of 402 months have a p value of 0.10 or less. After distributionmatching using the 20% initial draw setting, the KS statistic is 0.0585 (average p value = 0.35; 135 months have a p value = 0.10 or less); when the initial draw is 100 firms the average KS statistic is 0.0496 (average p value = 0.89; one month has a p value = 0.10 or less). Average crosssectional regression coefficients for all four IBES
CofE metrics in the unadjusted IBES
CofE sample are reliably negative (tstatistics between −2.96 and −5.03) and reliably different from 1 (tstatistics between −2.30 and −4.86). After distributionmatching using either 20% of the sample or 100 firms, average crosssectional regression coefficients are reliably positive (tstatistics between 2.20 and 2.91) and, with the exception of the
CT CofE metric, are not reliably different from 1. In summary, dropping the requirement of Value Line data increases the sample size, does not eliminate the nonrandomness of the resulting
CofE sample returns and does not systematically alter the associations of the
CofE metrics with realized returns in the unadjusted sample.
6 Conclusions
This paper proposes and illustrates a practical solution to a pervasive issue in empiricalarchival accounting research, namely, datarestricted samples that are nonrandomly drawn from the reference sample to which the researcher would like to generalize results. Paired with an objective of maximizing the number of observations with values for all variables (“complete cases”), these nonrandom samples are effectively dictated by data availability. We describe, validate, and illustrate a distributionmatching technique that can be used to align the distribution of a nonrandom estimation sample with that of a reference sample. The foundation for this approach is resampling from the datarestricted nonrandom subsample to minimize the distance between the marginal sample distribution and the marginal reference distribution.
We illustrate the effectiveness of the distributionmatching approach in tests of associations between returns and five popular implied cost of equity (
CofE) estimates. This setting is of interest in its own right, given the practical and theoretical importance of the association and the weak and mixed results in previous research. Our analysis shows that associations between realized returns and
CofE metrics are influenced by the properties of the realized returns distribution used to estimate the associations. Our reference sample is CRSP firms with at least 12 consecutive monthly returns during 1976–2009; our test sample is firms with sufficient data to calculate the
CofE measures. The latter sample is a substantially smaller, nonrandom subsample of the former. We first show that associations between realized returns and
CofE metrics are weak or negative, as in prior research. After distributionmatching, so that the resulting returns distribution mimics the returns distribution in the reference sample, we find reliably positive correlations between realized returns and most
CofE measures, as predicted by theory. This result suggests that several implied
CofE measures used in the accounting literature have greater construct validity than prior results suggest.
^{51} We also discuss two alternative approaches: multiple imputation (which performs well as a potential alternative to distributionmatching in our setting, albeit at the cost of additional assumptions) and selectiontype models (which do not perform well in our setting).
Viewed broadly, our analysis implies that nonrandomness of samples resulting from data requirements may lead to conclusions that do not generalize to a reference sample selected by the researcher. We demonstrate how to use available information about a marginal reference distribution of one variable of interest (in our setting, realized returns) to construct samples that mimic a reference distribution more closely than can an unmodified sample whose composition is dictated by data requirements. Highlighting an important caveat to the goal of maximizing the size of a dataconstrained research sample, our analysis suggests maximization of a dataconstrained sample may not be goalcongruent with increasing the generalizability from such a sample.
Our findings suggest researchers might benefit, in terms of increasing the generalizability of results, from examining the impact of data requirements on the empirical distribution of the test model variables, in particular the variable whose distribution is most affected by the availability of other variables of interest. We believe the approach we discuss, modified to suit the specific research context, will assist future research by providing an explanation for weak or counterintuitive results from datarestricted samples. Distributionmatching might also benefit future research by helping to coordinate across studies that address either similar questions using different samples. Comparisons of results across studies would be facilitated to the extent many researchers can define, construct and analyze a common reference sample.
Acknowledgements
We appreciate financial support from Duke University’s Fuqua School of Business, ESMT and the Frankfurt School of Finance and Management. We thank the Editor, Peter Easton, and two anonymous referees for their helpful guidance as well as Robert Bartels (MEAFA conference discussant), Mary Barth, Alex Belloni, Alon Brav, Federico Bugni, Judson Caskey, Qi Chen, Shuping Chen, Dain Donelson, Ron Dye, Jason Hall, Xu Jiang, Bill Kinney, Stephannie Larocque, Charles Lee (Stanford Summer Camp discussant), Fan Li, Xi Li (SMUSOAR Symposium discussant), John McInnis, Maria Ogneva (FARS Meeting discussant), Panos Patatoukas, Jerry Reiter, Hanna Setterberg, Yong Yu, and seminar participants at Dartmouth College, Stockholm School of Economics, University of Indiana, University of Iowa, University of Maryland, University of Munich, University of Notre Dame, University of Texas, Singapore Management University’s SOAR Symposium 2013, FARS Midyear Meeting 2014, the Stanford University Accounting Summer Camp, the 9th MEAFA Research Meeting at the University of Sydney, and the 5th International Corporate Governance Conference at Tsinghua University for their comments and suggestions. Early drafts of the paper were circulated under the title “Association Tests of Realized Returns and Risk Proxies Using NonRandom Samples.”
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Footnotes
1
We define nonrandomness by comparing a datarestricted sample to a specified reference sample to which the researcher would like to generalize results. Our reference sample is the population of CRSP firms with at least 12 consecutive monthly returns during February 1976 to July 2009 (the Full Returns sample).
3
The problem is equivalent to an “item nonresponse” in an otherwise complete questionnaire. The item is known to exist, but the data are not available to the researcher.
5
That is, we find a tradeoff between sample size/test power and generalizability. In contrast with the standard approach of using the largest possible number of observations with complete data on both variables, we show that it is not necessarily the case that datadictated samples of maximized size lead to unbiased inferences.
6
These regression coefficients are interpretable as implied factor premia. (For example, the coefficient in a regression of excess returns on market beta can be interpreted as the implied market risk premium.)
7
We acknowledge that research can, and sometimes should, be performed on restricted or even intentionally biased samples. In those cases, results are not intended to be generalizable to a reference sample. We also acknowledge that, if the researcher’s test sample is known to resemble the researcher’s reference sample with respect to the dependent variable, the issue we consider does not arise.
8
In the empirical example described later, data on
y (realized returns) are available for all firms in the reference sample while data on
x (
CofE metrics) are missing for 84% of observations in the average crosssection.
9
In this discussion, the subsequent simulations and most of the empirical work, we focus on the correlation coefficient not the regression coefficient because the former is not affected by changes in the (relative) standard deviations of the two variables. Therefore mechanical changes in standard deviations, for example, because the reference distribution is more dispersed or because of a reduction of the number of observations, will not confound our analysis. Examining correlation coefficients lets us demonstrate the effects of distributionmatching in isolation. We discuss the (equivalent) effects on the regression coefficients in Section
4.4.2.
10
Figure 2b contains visual evidence that this equality is maintained after distribution matching in the specific empirical example discussed later in the paper. The results show a very small difference in economic terms in averages for the Value Linebased
CofE metric before and after binbased distribution matching. The difference (visually) increases towards the extreme realized returns because of the paucity of observations in the tails. In none of the 101 returns bins is the difference significant at the 0.01 level (results not tabulated). Analogous differences using the other
CofE metrics are, if anything, generally smaller than the differences for the Value Linebased
CofE metric.
12
Hotdeck imputations similarly use the entire reference sample as a test sample by filling in the missing values in incomplete observations using realized values from “donor” observations that are similar to the “recipient” observations based on a proximity metric, usually measured using complete variables for both observations. While our approach also uses only realized values of the missing variable, we resample
whole observations from the restricted sample to match the known distribution of one variable and thereby preserve
pairs of the variables of interest. While distributionmatching might decrease the size of the test sample, hotdeck imputations (similar to multiple imputations) aim to maximize its size.
13
In simulations, we show that even minimal truncation can induce large bias in correlation coefficients in nonrandom samples of the outcome variable.
14
As the simulations illustrate, our procedure can also be used with a theoretically derived reference distribution.
15
An alternative is to oversample from selected groups to ensure the groups are surveyed in the first place. Subsequently, observations from the selected oversampled groups are assigned the (lower) population weight.
16
Briefly, those conditions are (1) the (largely untestable) assumption of bivariate normality of selection model and test model residuals and (2) the assumption that the selection model can be performed on a random sample of the reference sample. Many authors document the sensitivity of test results with respect to even minor departures from the normality assumption, leading to biases that may exceed the bias from standard completecase analyses. Due to this sensitivity, some authors go so far as to question the usefulness of selection models in practice (e.g., Enders
2010).
17
Details about the design of our simulations and the generation of three distinct nonrandom samples from the simulated population data are available from the corresponding author upon request.
18
Vuolteenaho (
2002) concludes that cash flow news is the main driver of firmspecific realized returns. Elton (
1999) observes there are periods exceeding 10 years during which realized stock returns are, on average, less than the riskfree rate, thereby questioning whether realized returns are a reasonable proxy for expected returns. He concludes that realized returns are “a very poor measure of the expected return,” although they continue to be used in asset pricing tests without so much as a “qualifying statement,” and suggests exploring exante cost of capital measures rather than realized returns.
19
The
CofE metrics are inferred from valuation models relating expectations of future cash flows, dividends or earnings to current price. By construction, these
CofE metrics are derived from “static” valuation models and therefore are not affected by “news” over a measurement period in the same way that realized returns might be affected.
20
Related work tries to increase the association between realized returns and the respective variable of interest by filtering out an expected (as opposed to nonexpected) return component. For example, Easton and Monahan (
2005) and Hecht and Vuolteenaho (
2006) use a variance decomposition approach to separate realized returns into expected return, cash flow news, and discount rate news components. Easton and Monahan use the components to explore the weak correlation between realized returns and implied cost of capital metrics. Hecht and Vuolteenaho use the components to explore the low correlation between realized returns and contemporaneous earnings.
21
In their firmspecific tests, one proposed method yields tstatistics between −0.52 and 1.93 for five implied cost of capital proxies and the other method yields tstatistics between −0.50 and 1.58.
23
Relative to a variance decomposition approach or a newspurging approach, we require no assumption about either the determinants of the expected return component or the functional form of the relation between news and returns. Relatedly, the measurement intervals of variables in an expected returns model do not dictate the data frequency in our tests, and disaggregated (e.g., monthly) data can be used.
24
The 12monthlyreturns requirement does not lead to a nonrandom returns sample. Across the 402 sample months, the average KS statistic comparing our reference sample with the CRSP returns universe is 0.0044 (average
p value = 0.86).
25
Because all tests are performed on monthspecific cross sections, using realized returns instead of excess returns yields equivalent regression and correlation coefficients. We use excess returns and do not further discuss raw returns.
28
The slope coefficient from a regression of realized excess returns on
CofE equals the correlation coefficient times the ratio of the standard deviation of the excess returns to the standard deviation of the
CofE estimate. Using the average results in Panel B of Table
2, this ratio ranges from 12.9 (VL
CofE) to over 25 (GLS
CofE). We use the correlation coefficient to capture the strength of association for two reasons. First, we wish to abstract from the effects of differing standard deviations across
CofE metrics. Second, our distributionmatching approach might affect the standard deviations of returns and
CofE metrics differently, inducing a change in the regression coefficient unrelated to the magnitude of the correlation. In Section
4.4.2, we report both correlation and regression coefficients.
29
While prior research has mostly used annual data, we use monthly versions of the
CofE estimates because asset pricing tests commonly use monthly returns.
30
For the timeseries regressions given by Equation (
7a), we use the more common specification with excess returns to estimate factor betas. As all association test results are averages from crosssectional estimations, using returns or excess returns is equivalent.
31
Prior research using firmspecific tests, as opposed to portfolio tests, also finds unexpected results for the
HML factor. For example, in their firmspecific tests in Table
4, Panel D, Core et al. (
2008) document a negative (sometimes weakly significant, sometimes insignificant) relation between the
HML factor beta and realized returns. Similarly, Gagliardini et al. (
2016) show a significantly negative
HML premium (their Tables
1 and
2). In portfolio designs (e.g., tests on size/booktomarket portfolio returns), the sign on the
HML factor betas is generally positive in prior literature.
32
The Random Subsample results in column 2 of Table
4 are based on averages of 20 random subsamples drawn from the Full Returns sample.
33
The nonparametric KS statistic captures any difference between two distributions, not limited to the first four moments.
34
As location of the distribution has no impact on either correlation coefficients or regression coefficients, we standardize both distributions (reference and current sample distribution) to a mean of zero before computing the KS statistic.
35
The
CofE sample contains on average 955 firms per month. Therefore 1000 iterations effectively allows each firm to enter the distributionmatched sample, to the extent its inclusion leads to a closer match to the returns distribution of the reference sample.
36
The KSbased distributionmatching approach can, in principle, be used to construct multiple subsamples, which can, in turn, be analyzed separately and then aggregated, resembling multiple imputation. The benefit of such an approach might include correctly specified crosssectional standard errors, which are of little interest in the FamaMacBeth design we use.
37
Although the design choices in this binbased approach are admittedly ad hoc, binbased sampling approaches are welldocumented as well as computationally more efficient.
38
This approach sharpens both goodnessoffit and poornessoffit in an unbiased fashion. That is, if a given bin in the
CofE sample contains realizedreturn/
CofE pairs that fit poorly, this approach will exacerbate that poor fit when the sampling percentage increases for that bin, and vice versa if the bin contains pairs that fit well. When sampling percentages are reduced, the opposite is the case.
39
The estimation on a random subsample will suffer from a loss in efficiency, compared to the estimation in the population, but results remain unbiased (as also shown in Table
4).
40
Lennox et al. (
2012) illustrate the sensitivity of even qualitative test results to selection model specification.
41
Firm age, as defined, and the factor betas are available for all observations. We use the log of all characteristics (firm age, size, volume, and booktomarket). We acknowledge that CRSP does not contain volume data for NASDAQ firms prior to November 1982; therefore sample losses for selection models that include volume are largely due to that earlier period, while coverage afterward is almost complete.
42
As a second test of the effectiveness of including an inverse Mills ratio, we use a similar adjustment in the factor beta regressions for the actual
CofE sample, aiming to restore factor premia obtained from the Full Returns sample (results not tabulated). Specifically, we use the variables in selection Model IV and rerun the crosssectional asset pricing tests using a factor beta and the inverse Mills ratio. When we include the inverse Mills ratio, factor premia estimates from the
CofE sample are hardly affected (differences range from −0.0004 to −0.0001), insignificant at conventional levels, and qualitatively different from the Full Returns sample results.
43
Standard statistical software packages like SAS and Stata include commands for performing multiple imputations, for diagnostics to check for the convergence of the estimation, and for the aggregation of the test results from the imputed datasets. We used the MI procedure and the MIANALYZE procedure in SAS for these functions. We used the expectation maximization (EM) algorithm to determine the distribution of possible parameter values for the imputation of the
CofE metrics, using the complete excess returns data. Using this solution as a starting point, we use an iterative Markovchain Monte Carlo approach to draw from that distribution and construct 10 datasets, each of the size of the Full Returns sample, with the same (complete) excess returns data and a full vector of the
CofE metrics consisting of measured and imputed values. The 10 datasets can be analyzed independently and results aggregated. We formulate a monthspecific imputation model for each
CofE metric using only the (complete) excess returns data. To improve the model fit, we split each monthly crosssection at the median, into terciles and into quintiles, resulting in two, three or five groups, allowing for different intercepts and coefficients in each group.
44
Random inspection suggests that the incidence of negative imputed
CofE values is small in the average crosssection.
45
The Execucomp database covers S&P 1500 firms since 1994, but other compensation data sources can be more restrictive. See, for example, Brookman et al. (
2006) for an overview.
46
Average mean excess return, standard deviation and skewness differ between the subsamples. Specifically, the subsamples that do not meet the sample selection criteria have larger average mean excess returns, larger standard deviations of excess returns, and greater positive skewness of excess returns (results not tabulated).
47
Recall that the
HML beta is negative in our firmspecific setting, consistent with other studies using firmspecific returns (e.g., Gagliardini et al.
2016). Consequently, its tstatistic becomes more negative as the proportion of positive returns increases.
48
We thank an anonymous referee for suggesting this test, which provides an opportunity to apply distribution matching in an annual returns setting, to complement the monthly returns setting analyzed in most of this paper. We believe there is no
ex ante reason to predict whether the datarestrictions in the Richardson et al. paper (including requiring certain accounting data) will or will not be consequential in terms of affecting the distribution of returns for the filtered (datarequirementsconstrained) sample.
49
Our initial sample is about 37% larger than Richardson et al.’s, regardless of whether we use the current version of Compustat or a legacy version intended to approximate the version available in the early 2000s. Eliminating lowpriced stocks (price less than $5 at the end of Month +3 after the fiscal yearend) results in a sample whose size is similar to that in Richardson et al. Therefore we discuss (untabulated) results using this pricefiltered returns sample as the reference sample of returns as our main results. We obtain qualitatively similar results using the larger, unfilteredreturns reference sample (i.e., a sample analogous to the sample in the monthly returns setting analyzed in this paper).
50
When we impose data requirements separately for each of four IBESbased
CofE metrics and the VL metric, the monthly average number of observations are: 2130.7 (
CT CofE); 2169.7 (
OJN CofE); 2063.1 (
MPEG CofE); 2226.0 (
GLS CofE); and 1495.4 (
VL CofE). In all five cases, the KS statistic rejects at the 0.0016 level or better the hypothesis that the realized returns distributions of the
CofE samples are similar to the realized returns distribution of the reference sample.
51
We emphasize that we implement the
CofE metrics as originally developed. The fact that the metrics are positively correlated with realized returns in our distributionmatched sample does not mean they cannot be improved upon, either by developing new metrics altogether, by adjusting input variables, or by developing alternative empirical implementations of these metrics. For a thorough discussion and analysis, see Easton (
2007).