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Erschienen in: International Tax and Public Finance 2/2022

18.06.2021

Gini and undercoverage at the upper tail: a simple approximation

verfasst von: Pablo Gutiérrez Cubillos

Erschienen in: International Tax and Public Finance | Ausgabe 2/2022

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Abstract

This paper investigates the impact of top-distribution undercoverage on the Gini coefficient. First, we show that failing to correct for underreporting and nonresponse at the top does not necessarily result in an underestimated Gini coefficient. Then, we establish analytical conditions under which the Gini coefficient of the uncorrected distribution is higher than the Gini coefficient of the true distribution, i.e. the distribution that incorporates underreporting and/or nonresponse. In addition, we propose a Gini approximation based on the Atkinson approximation \(G=G_{1-p}\cdot (1-S_{p})+S_{p}\) to correct for underreporting at the top. Under plausible assumptions, the approximation proposed is very close to the real Gini coefficient. We also show that before correcting for underreporting we need to first address the issue of nonresponse as otherwise the approximation may be strongly upper biased. Finally, this work proposes a procedure to estimate the fraction of nonrespondents at the very top under the assumption that underreported incomes belong to individuals with incomes higher than those reported in a household survey. To evaluate the proposed methodologies, this paper uses Chile and Canada as examples, where we include undistributed business profits—an underreported source of income—to measure income inequality.

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Fußnoten
1
The Gini coefficient is not the unique index used for measuring income inequality. For a discussion about the comparison of two income distributions and inequality indexes see Atkinson (1970).
 
2
For a broader discussion about the limitations of the Gini index, see Cowell and Flachaire (2018), Alvaredo et al. (2017), and Osberg (2017). For elegant usages of this index, see Corvalan (2014) and Modalsli (2017), and for an introduction to the Gini coefficient, see Ceriani and Verme (2012).
 
3
Undercoverage not only exists at the top of the distribution: Higgins et al. (2018) and Bollinger et al. (2018) show that the entire distribution faces undercoverage, and Ceriani and Verme (2019) show that “inequality almost invariably increases by adding observations on the tails of an income distribution but that missing a few observations at the top is much more relevant than missing many observations at the bottom.”
 
4
In the statistical literature, the term nonresponse can be used as unit nonresponse or item nonresponse (Lustig 2019). Here, we refer as nonresponse as unit nonresponse, that is, when there are missing individuals instead of missing data.
 
5
Flores (2021) also shows that household surveys do not adequately account for capital income. This problem is also considered underreporting.
 
6
There are other biases that may affect measurement of top income inequality. For instance, small sample bias (Taleb and Douady 2015), that is, incomes at the very top span very large dollar ranges—the difference between the top and bottom of the top 0.01% is measured in multiple millions. But only a few hundred dollars separate people whose incomes are near the median—e.g. within the 51st percentile. When a random sample is drawn, sampling variability can matter at the top end even if it is inconsequential for middle incomes.
 
7
Diaz-Bazan (2015) present another exact interpolation which is more exact than the Atkinson approximation, but requires additional data.
 
8
As was discussed by Cowell and Flachaire (2015) and Higgins et al. (2018) estimating this p is a major challenge.
 
9
Indeed, as Bollinger (1998) states “there is little or no relationship between the probability of giving a correct response and the level of income. The finding that (measurement) error is negatively correlated with true income is true only for “gross errors” nor for the probability of giving an accurate response.”
 
10
The most realistic case is that of random undercoverage, i.e. when the probability may increase with income. For example, this issue is broadly studied by Blanchet et al. (2019).
 
11
There might be some incentives to underreport income for someone that belongs to the bottom part of the distribution. For example, some refundable tax credits are maximized when someone works up to a threshold, after that threshold, the tax benefit decreases with income. For example, the Earned Income Tax Credit (EITC) in the US or the Canada Workers Benefit (CWB).
 
12
This assumption is consistent with the measurement error literature (Bollinger 1998; Bound and Krueger 1991; Abowd and Stinson 2013)
 
13
Bollinger et al. (2018) present evidence that the income distribution, including nonrespondents, is more unequal than the observed distribution.
 
14
We can easily prove this by applying the population principle.
 
15
\(\phi\) is greater than z because \(\phi\) also includes the proportion of underreporters captured by the household survey. One way to know \(\phi\) and z is by having two harmonized income-distribution information sources: a household survey and income-tax data. Harmonized sources are essential, but in practice, such harmonized sources may not exist—e.g., Atkinson et al. (2011) used nonharmonized data, but Jenkins (2017), Burkhauser et al. (2018), and Piketty, Saez, and Zucman (2018) did.
 
16
Results are presented via a single parametric distribution. However, the bottom part of the distribution could also be approximated with Singh-Maddala, Dagum, or GB2 distributions, which produce similar results. Indeed, as Proposition 5 shows, the bias of the underreporting approximation does not depend on the distribution of the bottom p. However, by Proposition 6, a more equal distribution at the bottom p reduces the bias of the nonresponse approximation.
 
17
For simplicity, results are presented using a Type I Pareto distribution; however, as Atkinson (2017) shows, Pareto type I is not a perfect tool for studying top income shares and is rather “at best a convenient first summary of the extent of the income concentration.” In addition, Jenkins (2017) showed (for the United Kingdom) that a Pareto type II is preferable to the Pareto type I typically used at the thresholds. Blanchet et al. (2018) also used a Pareto type II to fit income inequality at the top.
 
18
Note that the Gini coefficient of \((x^{*},\ldots ,x^{*}\)) is 0.
 
19
The Gini coefficient of the observed top 1% and the nonrespondents is \(G_{1\%+\theta }=\frac{S_{1\%}}{S_{1\%}+S_{\theta }}\frac{0.01}{0.01+\theta }G_{1}^{*}+\frac{S_{\theta }}{S_{1\%}+S_{\theta }}\frac{\theta }{0.01+\theta }G_{\theta }+\frac{S_{\theta }}{S_{1\%}+S_{\theta }}-\frac{p}{0.01+\theta }\).
 
20
Tax planning using small businesses in Canada and Chile was broadly studied in the literature. See for instance Fairfield and Jorratt De Luis (2016), López et al. (2016) and Wolfson et al. (2016).
 
21
For more details related to the empirical application see Tables 2 and 3 in the Appendix.
 
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Metadaten
Titel
Gini and undercoverage at the upper tail: a simple approximation
verfasst von
Pablo Gutiérrez Cubillos
Publikationsdatum
18.06.2021
Verlag
Springer US
Erschienen in
International Tax and Public Finance / Ausgabe 2/2022
Print ISSN: 0927-5940
Elektronische ISSN: 1573-6970
DOI
https://doi.org/10.1007/s10797-021-09671-4

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