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Managerial risk incentives and accounting conservatism

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Abstract

We provide empirical evidence of the effect of managerial risk incentives on financial reporting conservatism. We hypothesize that firms use greater accounting conservatism as a means of addressing increased firm risk arising from excessive managerial risk incentives provided by option compensation. Consistent with this hypothesis, we find a positive association between excessive managerial risk incentives and accounting conservatism measured as asymmetric timeliness of loss recognition. By contrast, we find no impact by normal (anticipated) risk-taking on accounting conservatism. Further analysis shows that the association between excessive managerial risk incentives and accounting conservatism is more pronounced when firms face more severe debtholder–shareholder conflicts. We also find that while cost of debt financing is positively associated with both anticipated and excessive risk incentives, the relationship with the latter is weakened by timelier loss recognition, suggesting firms with heightened risk incentives could economically benefit from using more conservative accounting.

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Notes

  1. The maximum amount of payment received by debtholders is limited to the face value of the debt. Debtholders receive less than the contracted sum when the firm can’t produce enough net assets to cover the promised payments.

  2. We focus on the option component of equity compensation because Guay (1999) finds that convexity provided by common stock is small in magnitude and of little economic significance relative to that of stock options. Our inferences remain unchanged when we include common stockholdings in the calculation of vega.

  3. While the focus of this study is excessive risk taking incentives, we also examine the effect of normal, or anticipated, risk taking in Sect. 4.5.

  4. Ball and Shivakumar (2005) argue that it is conditional conservatism that constrains opportunistic managerial behavior such as wealth transferring, investment in value-reducing projects, and therefore increases the efficiency of debt contracts. Unconditional conservatism, defined as an accounting bias towards reporting low book values of stockholder equity, on the other hand, seems ineffective or at best neutral in enhancing contracting efficiency.

  5. Prior research also suggests other mechanisms such as making bond indenture more restrictive (Jensen and Meckling, 1976), adding security provisions (Stulz and Johnson 1985), and using short-term debts (Brockman et al. 2015). The use of these alternatives would work against finding results supporting our hypothesis.

  6. Conservatism has influenced accounting practice for centuries. Studies have shown that accounting practice has become more conservatism in recent years (Basu 1997; Ball et al. 2000; Holthausen and Watts 2001). Lobo and Zhou (2006) documents an increase in conservatism in financial reporting following the Sarbanes–Oxley Act of 2002.

  7. Ball and Shivakumar (2005) suggest that managers who know ex ante that investment losses will be recognized during their tenure, “are less likely to make negative net present value investments, such as ‘pet’ projects or ‘trophy’ acquisitions” (p. 87).

  8. Debtholders may demand conservative reporting to protect themselves from normal managerial risk-taking. However, the effectiveness of conservatism neither uncertain ex ante nor measureable ex post compared to more direct risk-adjusting mechanisms such as high yield and/or restrictive covenants.

  9. The stockholders’ demand for conservatism is a function of the alignment effects stemming from compensation wealth incentives and managerial risk incentives. Prior studies show that the alignment effect arising from risk incentives is second order to that from compensation wealth incentives (e.g. Core et al. 2003; Hayes et al. 2012). We control for stockholders’ demand for conservatism in all our regression analyses.

  10. Prior research also suggests other mechanisms such as making bond indenture more restrictive (Jensen and Meckling 1976), adding security provisions (Stulz and Johnson 1985), and using short-term debts (Brockman et al. 2015). The use of these alternatives would work against finding results supporting our hypothesis.

  11. Debt contracting is the primary explanation for conservative accounting (e.g.,Watts 2003a). Prior literature finds that accounting conservatism decreases debt-contracting costs. Ahmed et al. (2002) and Zhang (2008) find that conservatism is associated with a lower cost of debt.

  12. Litigation risk (LIT) is measured as the probability of litigation estimated using the coefficients from the litigation risk model of Kim and Skinner (2012), model 2 in Table 7. Alternatively, we define litigation risk as a dummy variable which takes the value of 1 if firm is in a high litigation risk industry as defined by 4-digit SIC codes 2833–2836, 3570–3577, 7370–7373, 3600–3674, and 5200–5961; or 0 otherwise. Our results are qualitatively similar using these different definitions.

  13. We further address the issue of endogeneity by performing a changes analysis and examining the change in accounting conservatism in response to the exogenous shock from FAS 123R in Sect. 5.1.

  14. Leverage is used by Brockman et al. (2015) as proxy for debtholder-stockholder conflict. Though commonly used to measure the degree of agency conflict associated with debt, leverage has been reported by prior studies to capture many other firm characteristics. For example, Lang et al. (1996) show that leverage proxies for growth opportunities, and Press and Weintrop (1990) use leverage as proxy for accounting based constraints. To address the concern, we use other proxies of debtholder-stockholder conflicts, as suggested by Smith and Warner (1979) in the study.

  15. Unlike the traditional Black–Scholes–Merton model, which assumes risk-neutrality where all assets are expected to growth at risk-free rate, the derivation of the probability of default is a function of the actual distribution of future assets value, which is a function of the expected growth in asset’s value, µ. Following Hillegeist et al. (2004), we calculate expected growth rate in asset’s value, µ, as the larger of actual asset growth rate and risk free rate. We set the risk-free rate to the one-year treasury bill rate.

  16. Previous studies on bankruptcy prediction frequently use composite measures estimated with a set of accounting variables (ratios), with Altman’s (1968) Z-Score and Ohlson (1980) O-Score being the most commonly used ones. The market-based probability measure outperforms these accounting-based measures in several ways. First, accounting-based measures are inherently backward looking and thus are not informative about the future status of the firm. The option-based probability of default measure, on the other hand, relies on market-based variables and aggregates information from other sources in addition to the financial statements, hence having more predictive power. Second, accounting-based measures generally fail to incorporate asset volatility. As noted by Hillegeist et al. (2004), asset volatility is a crucial variable in bankruptcy prediction in that it captures the likelihood of the firm being unable to repay its debts due to declines in asset value. Finally, total liability (and therefore leverage) is an important component of nearly all the accounting-based measures. As a result, these accounting-based measures are highly correlated with firm leverage, which serves as our first proxy for debtholder-stockholder conflict. For example, the correlation between leverage ratio and O-score for the sample used in this study is 0.562, while the correlation between leverage ratio and DTD is only 0.153. Given these advantages, we employ DTD as the primary measure of financial distress. We conduct robustness checks using Ohlson’s (1980). Our results are not sensitive to the choice of bankruptcy prediction measures.

  17. The dollar value of VEGA is highly skewed and thus justifies the log transformation.

  18. Contrary to the findings by Guay (1999), we find VEGA to be negatively related to investment expenditure (INVEXP).

  19. FAS 123R became effective on June 15, 2005. Results are qualitatively similar if we widen sample range to include years between 2002 and 2007.

  20. Guay (1999) finds that the vega of stock compensation is negligible. Therefore a switch from the option-based to stock-based compensation results in a monotonous decrease in option vega.

  21. The coefficient on RES_VEGA*D*R is 0.063 (p value < 0.01) for the subsample with the above-median dividend ratio and − 0.020 (p value = 0.839) for the below-median group, with the difference significant at the 0.01 level. The coefficients on RES_VEGA*D*R are 0.062 (p value < 0.01) and 0.013 (p value = 0.200) for subsamples with the above- and below-median DTD values, respectively, with the difference significant at the 0.05 level.

  22. The Securities and Exchange Commission (SEC) requires detailed disclosure of CEO pensions and deferred compensation, including the actuarial net present value and annual increases of such balances, effective the 2006 fiscal year.

  23. The coefficients on RES_VEGA*D*R are − 0.069 (t value = − 2.50), − 0.071 (t value = − 2.54), − 0.068 (t value = − 2.38), − 0.066 (t value = − 1.25), and − 0.067(t value = − 2.41), respectively, when measures of payout-related, investment-related, financing-related, accounting-related, and overall covenants are included in Model (3).

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Authors and Affiliations

  1. The State University of New York – Canton, 34 Cornell Drive, Canton, NY, 13617, USA

    Chengru Hu

  2. California State University – Fullerton, 800 N. State College Blvd. SGMH - 4389, Fullerton, CA, 92831-3599, USA

    Wei Jiang

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  1. Chengru Hu
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Hu, C., Jiang, W. Managerial risk incentives and accounting conservatism. Rev Quant Finan Acc 52, 781–813 (2019). https://doi.org/10.1007/s11156-018-0726-5

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