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Review of Accounting Studies

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01-03-2009

Accounting conservatism and corporate governance

Authors: Juan Manuel García Lara, Beatriz García Osma, Fernando Penalva

Published in: Review of Accounting Studies | Issue 1/2009

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Abstract

We predict that firms with stronger corporate governance will exhibit a higher degree of accounting conservatism. Governance level is assessed using a composite measure that incorporates several internal and external characteristics. Consistent with our prediction, strong governance firms show significantly higher levels of conditional accounting conservatism. Our tests take into account the endogenous nature of corporate governance, and the results are robust to the use of several measures of conservatism (market-based and nonmarket-based). Our evidence is consistent with the direction of causality flowing from governance to conservatism, and not vice versa, indicating that governance and conservatism are not substitutes. Finally, we study the impact of earnings discretion on the sensitivity of earnings to bad news across governance structures. We find that, on average, strong-governance firms appear to use discretionary accruals to inform investors about bad news in a timelier manner.

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Following Beaver and Ryan (2005), we refer to this news-dependent conservatism as conditional. Other authors label it as ex post conservatism, income statement conservatism, or earnings conservatism. Unconditional or news-independent conservatism—also labeled ex ante or balance-sheet conservatism—in turn, refers to the persistent understatement of shareholders’ equity that results from historic cost accounting and underrecognition of certain intangible assets due to the accounting rules (Feltham and Ohlson 1995). In the paper, we only focus on conditional conservatism as it plays a clear role in the contracting and monitoring functions of corporate governance. However, it is difficult to see how contracting is affected by conservatism in the form of an unconditional accounting bias of known magnitude. Rational agents would simply invert the bias. If the bias is unknown, it can only reduce contracting efficiency (Ball and Shivakumar 2005).

Watts (2003b) argues that tax and regulation also contribute to conservatism; however, the empirical evidence thus far offers more limited evidence on the contribution of these factors to conservatism.

Our use of the expression strong (weak) governance is purely descriptive. It is not intended to mean that strong governance is better than weak governance.

For example, internal governance mechanisms such as independent boards of directors and audit committees have been shown to constrain aggressive practices, limiting the incidence of income-increasing earnings management (Beasley 1996; Klein 2002; Peasnell et al. 2005). Similarly, recent research shows that independent audit committees hire better quality auditors (Abbot et al. 2003) that, in turn, impose more conservative accounting choices (Basu et al. 2001; Chung et al. 2003).

Literature on this field provides mounting evidence that efficient corporate governance results in lower agency costs and that internal and external governance structures are associated to firm performance. For example, Cremers and Nair (2005) show that firms with strong external and internal governance generate abnormal returns of 10% to 15%. Core et al. (1999) find that less effective boards of directors—characterized by the CEO holding the chairman position; larger size; directors appointed by the CEO; and the presence of gray outside directors, old directors, and busy directors—are correlated with higher levels of CEO compensation after controlling for economic determinants of compensation; moreover, they find that predicted excess compensation, based on the governance structure of the firm, is negatively correlated with stock returns 1, 3, and 5 years ahead.

Gompers et al. (2003) examine 24 provisions: anti-greenmail, blank-check preferred stock, business combination laws, bylaw and charter amendment limitations, classified board, compensation plans with change in control provisions, director-indemnification contracts, control share cash-out laws, cumulative voting requirements, director’s duties, fair-price requirements, golden parachutes, director indemnification, limitations on director liability, pension parachutes, poison pills, secret ballots, executive severance agreements, silver parachutes, special meeting requirements, supermajority requirements, unequal voting rights, and limitations on action by written consent.

Like Bertrand and Mullainathan (2001), we use unit weights to construct Totgov following the recommendations of Grice and Harris (1998), who find that unit-weighted composites exhibit better psychometric properties than alternative weighting schemes.

Prior studies (Givoly et al. 2007; Callen et al. 2006) express their distrust of inferences drawn from the Basu (1997) model if used in a time-series (firm-specific) approach. We use a cross-sectional approach.

Ryan (2006, Footnote 2) states that “two well-known empirical results together imply the biases identified by Dietrich et al. are likely to be fairly small and so biases in returns-based measures of asymmetric timeliness are likely to be correspondingly small. First, the low R2s observed in contemporaneous returns-earnings regressions suggest that the extent to which earnings causes returns is tiny compared to the extent to which both variables are determined by other, more primitive information. Second, a large literature, only some of which employs the reverse regressions of earnings on returns used to estimate asymmetric timeliness, exists that shows returns typically reflect information on a timelier basis than earnings.”

Basu uses the annual stock rate of return measured from 9 months before fiscal year end t to 3 months after fiscal year-end t. However, most subsequent studies use the fiscal year. Measuring returns 3 months after fiscal year-end is aimed at giving time to the market to incorporate information in contemporaneous earnings. Using fiscal year returns avoids returns being distorted by new information (different from earnings) coming to the market. Our results are not affected by this choice.

The inclusion of additional control variables such as the incidence of losses and earnings variability (Francis et al. 2004) does not change the inferences. Neither does including as a proxy for growth opportunities, the book-to-market value of assets ratio. We exclude this last variable because it also captures a certain degree of conservatism.

We are grateful to an anonymous referee for this insight.

Managers may also manipulate the timing and level of cash flows (e.g., Roychowdhury 2006; Bushee 1998; Bartov 1993), however, due to its low flexibility and high visibility, this is expected to be a residual form of earnings management (Peasnell et al. 2000).

Our data covers the period 1992 through 2003. The IRRC data is only available for 1990, 1993, 1995, 1998, 2000, and 2002. Gompers et al. (2003) report that for the majority of firms there is little time-series variation in the index. Taking advantage of this fact, like Cremers and Nair (2005), we align the index values available for 1990 with firm data for 1992, the index values for 1993 with firm data for 1993 and 1994, the index values for 1995 with firm data for 1995, 1996, and 1997, the index values for 1998 with firm data for 1998 and 1999, the index values for 2000 with firm data for 2000 and 2001, and the index values for 2002 with firm data for 2002 and 2003.

For parsimony, we only report the results that use the modified Jones model of Dechow et al. (1995) to estimate discretionary accruals. The results are not affected by the choice of accruals estimation method.

Fama and MacBeth (1973) regressions should be interpreted with caution. Basu (1999) gives a number of reasons against the use of mean annual regressions, related mainly to the parameters not being stationary.

Our estimate of discretionary accruals is based on the modified Jones model. This model only controls for two simple relations: between accruals and sales and accruals and property, plant, and equipment. This model would rarely capture other possible drivers of conservatism such as special items (restructuring charges and other one-time items). Managers also may use special items to affect conservatism. To assess this possibility, we augment earnings and the discretionary accruals estimate by adding the special items (Compustat item #17) deflated by beginning-of-the-period market value of equity. Then, we repeat the tests in Panel A of Table 2. Untabulated results indicate that the inferences still hold.

We also repeated this test including an industry × year interaction term and obtained the same inferences.

Notice that our proxies for conservatism—Basu’s (1997) earnings asymmetric timeliness, Ball and Shivakumar’s (2005) accruals asymmetric timeliness, and Givoly and Hayn’s (2000) average accruals—are different from the measure for the relevance of accounting numbers used by Bushman et al. Their measure captures earnings symmetric timeliness, which is closer to what the literature refers to as relevance.

In these tests we are unable to use the Heckman procedure as described in Sect. 3.3. The reason is that here we are partitioning the sample into strong and weak governance firms, and the probit regression that models governance choice cannot be applied to each partition separately. Nevertheless, all previous evidence indicates that the results are not biased by not taking into account the endogeneity of governance choice.

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Title: Accounting conservatism and corporate governance
Authors: Juan Manuel García Lara
Beatriz García Osma
Fernando Penalva
Publication date: 01-03-2009
Publisher: Springer US
Published in: Review of Accounting Studies / Issue 1/2009
Print ISSN: 1380-6653
Electronic ISSN: 1573-7136
DOI: https://doi.org/10.1007/s11142-007-9060-1

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