Debt Contracting and Information Shock: The Effect of Bankers on Accounting Conservatism

Transcription

1 Debt Contracting and Information Shock: The Effect of Bankers on Accounting Conservatism Abstract Using an informational shock to the information environment, this paper examines how non-affiliated financial executives (bankers) on boards relate to the level of accounting conservatism. We exploit IFRS mandatory adoption as an exogenous shock that might exacerbate the debtholders-shareholders conflict due to the wider use of fair value to analyze bankers role on accounting conservatism. Our results show that a significant increase in the level of conservatism for firms with banker representation follows the information shock/mandatory IFRS adoption. No such effect is observed if banker is not on board. These findings hold after considering other country-level factors that might play a role in shaping the demand of conservatism. In further analysis we find that the effect of bankers on the change in accounting conservatism is conditional on firm-specific incentives. 1. Introduction Accounting research has extensively analyzed the effect of monitoring oriented board on earnings quality (Klein, 2002; Vafeas, 2005; Peasnell et al., 2005; Krishnan et al., 2008). Board of directors, seen as a control mechanism, reduces agency conflict by improving the quality of accounting information. Garcia Lara et al. (2009) expand this view showing that monitoring oriented governance complements accounting conservatism in reducing conflicts among all contractual parties. However, extant literature (Hillman et al. 2000; Adams and Ferreira, 2007; Coles at al., 2011) has recognized that boards fulfill not only a monitoring role but also provides important connections and expert advice to top management team. Social network theory has emphasized the importance of interlocks as information intermediaries. In this vein, Shi et al. (2013) and Chiu et al. (2013) find that firms that share board members use the same accounting practices. Moreover, Coles at al. (2011) suggest that interlocking represents the derived demand for director service (emphasis in the original text; p. 2), thus suggesting, in line with Booth and 1

2 Deli (1999) and Byrd and Mizruchi (2005) that bankers are appointed as directors because of their debt- market related expertise. Building on social network theory, this paper analyzes the role of non-affiliated financial executives 1 (henceforth, bankers) on financial reporting outcome, and specifically on conservatism 2. We expect that bankers affect the level of conservatism, because: 1) conservatism has been extensively considered an efficient mechanism to mitigate debtholders-shareholders conflicts (Watts 2003a; 2003b); 2)bankers possess information about the conservatism demand arising from the debt market, which becomes an informational input into the accounting decision of the firm where they serve on boards. Moreover, the fact that bankers have first-hand information and debt-market expertise (Booth and Deli, 1999; Byrd and Mizruchi, 2005) may make other board members more comfortable with accepting such costly accounting choice. To analyze the role of bankers on accounting information we exploit an exogenous shock to the debtholders information environment given by the IFRS mandatory adoption. More specifically, the wide use of fair value enacted by the IFRS adoption represents a major breakthrough in financial reporting which triggers a change in the information environment and in firms level of conservatism. Indeed, IFRS amplifies the use of fair value, but it has many shortcomings for debt contract purposes. First, fair value might just reflect market liquidity rather than the fundamental 1 A board interlock occurs when a director of one firm sits on the board of directors of another firm (Mizruchi, 1989). When the shared director is an executive of a financial institution (banker) there is a financial interlock. If the banker sits on the board of an industrial company with which its financial institution has a lending relationship then the banker is considered affiliated, otherwise the bankers is considered a non-affiliated. In this study we focus only on non-affiliated bankers. 2 When studying the association between corporate governance and accounting conservatism, we treat governance as exogenous. Our approach is the same as that of Core et al. (1999, footnote 2), where they observe that, Following most prior empirical research in this area, we treat the board and ownership structures as exogenous, when economic theory would argue that these variables are endogenous. This well-established approach of treating governance structures as exogenous is reasonable, in the sense that some institutional features of contracting cause governance characteristics to be sticky. For example, directors serve for fixed terms, so naturally it takes time to change board members to adjust to a changed operating environment. Consistent with many prior studies, we argue that it is difficult for firms to have optimal governance structures at all times (Larcker et al., 2006). This assumption is more likely to hold in a difference-in-difference design to the extent that we are able to clean up the unobservable heterogeneity and the time invariant selection bias. 2

3 value of the assets. Second, fair value estimation is plagued by a lack of verifiability, especially when fair value is estimated using internal models and not market prices in active markets for identical assets. Third, fair value changes might originate in shock to both cash flows and discount rates that might be irrelevant for long-term debtholders because they both can reverse before debt maturity. For all these reasons, the IFRS trigger an information shock that may exacerbate the conflict between debtholders and shareholders thus increasing the need of conservatism, all else being equal. However, under IFRS the fair value model is either just allowed or the preferred accounting model but not the exclusive accounting model, with the only exception of equity instruments. Moreover, IFRS substitute unconditional conservatism with conditional conservatism, i.e. goodwill and development costs. Therefore, since IFRS do not fully determine the level of accounting conservatism, firms specific choices and then bankers are of extreme importance. To this end a banker may provide firms with first-hand information and direct knowledge about the demand of conservatism coming from the debt market. Therefore, we therefore predict a positive association between the presence of bankers on board and the change of conservatism as a consequence of the informational shock. The underlying intuition is that the average level of conservatism, the new equilibrium, in firms with bankers is higher than in firms without bankers in the post IFRS adoption period.. We empirically test our prediction across 14 EU countries by estimating annual panel regressions for IFRS adopter firms (treatment sample) and propensity matched non-ifrs adopter firms (control sample). We estimate yearly panel regressions and include fixed effects for industry, 3

4 country and three separate yearly time trends for Non-IFRS firms, IFRS firms with bankers and IFRS firms without bankers. Allowing for three flexible time trends should absorb unrelated shocks to yearly level of conservatism within the three firms groups. In addition, the use of a propensity score matching to identify the control group allows us to account for differences between treated and untreated firms while estimating the information environment shock effect. The combination of differences-in-differences estimation and propensity score matching has been widely used in empirical micro-economics (Heckman and Vytlacil 2007) as it alleviates concerns regarding the assumption of parallel patterns in the counterfactual scenario of no informational shock in a difference in difference research design. This approach strengthens our identification strategy. We start our analysis with an examination of the average effect of IFRS adoption on the level of conservatism measured as the C_SCORE metric, developed by Khan and Watts (2009). We find that the mandatory IFRS adoption does have a significant effect on the level of conservatism. However, this result disappears once we exploit firm-level heterogeneity in the presence of banker. We document that only firms with banker on board experience a significant increase in the level of conservatism after mandatory IFRS adoption. These findings hold after considering other country-level factors and firm-specific incentives that might play a role in shaping the demand of conservatism. Specifically, we use the following country-level factors: i) shock in accounting conservatism and ii) importance of debt market. The idea behind these additional tests is to illustrate that firm-level conservatism is induced by the presence of bankers rather than the change in accounting standard per se or the importance of the debt-market. Finally, to rule out the concern that results might be driven by firm-specific characteristics we perform a battery of 4

5 additional tests considering the intensity of the conflict between debtholders and shareholders over dividend policy and incentives to increase the level of accounting conservatism. Our study contributes to the literature in several ways. First, we contribute to the scant literature on corporate governance and conservatism. Given the importance of conservatism on debt contracting it is importance to understand the effect of a debtholder-oriented board on conservatism. To the best of our knowledge this is the first study that directly links non-affiliated bankers and conservatism 3. While a large literature in sociology, finance and management has examined the strategic and financial consequences of bankers on board, accounting implications of having a banker has been almost un-explored. We contribute to the literature on the role of bankers on board showing their role w.r.t financial reporting outcome. Additionally, we show that an exogenous shock to the information environment may alter the demand and under specific circumstances the choice of conservatism. Second, we contribute to the extent of literature on mandatory IFRS adoption, showing that the IFRS effects are heterogeneous in that they depend on the corporate governance. Third, there is a wide literature that examines the role of bankers on debt-ratio and firms financial policy. Overall, the findings confirm that bankers positively affect firms debt ratio. This paper contributes to this literature showing that the positive link between bankers and debt ratio may go through firms accounting policy. Particularly, by affecting the level of conservatism the presence of bankers may help firms to secure new debt. Our results must be interpreted with due regard to their limitations and to the caveats, discussed in more detail throughout the study. The rest of the paper is organized as follows. In section 2, we review the literature and develop our hypothesis. In Section 3, we outline the research design and describe the sample selection. 3 There is a concurrent paper that investigates the role of affiliated bankers on conservatism (Erkens et al., 2012). We discuss the differences in the Hypothesis section. 5

6 Section 4 discusses the main results. Section 5 discusses the results of additional analyses, and Section 6 concludes. 2. Related Literature and Prediction This paper builds on three streams of literature: conservatism, the role of bankers on board and the mandatory IFRS adoption. 2.1 Conservatism Conditional conservatism is commonly defined as a prudent reaction to uncertainty that imposes stronger verification requirements for the recognition of economic gains than for the recognition of economic losses, thus leading to earnings that reflect bad news in a timelier fashion than good news (Watts, 2003a, 2003b). In the debt-contracting explanation, conservatism occurs because it reduces agency costs associated with: 1) information asymmetries and loss functions among the contracting parties; and 2) the inability to verify the more informed parties private information (Ahmed et al., 2002; LaFond and Watts, 2008). 2.2 Conservatism and IFRS The framework issued by the International Accounting Standard Board (IASB) consider shareholders as the privileged users of accounting information and, in line with the equity valuation role of financial accounting, IFRS allows for a much more extensive use of fair value than any previous accounting regimes. The underlying assumption is that fair value represents in many circumstances a timelier and more informative estimation of future cash flows. As Ball et al. (2012) suggest there are many ways in which fair value is detrimental for the debtholders information set. First, many important IFRS standards (IAS 39; IAS 16; IAS 40; 6

7 IAS 41) require unrealized gains and losses due to changes in fair value of trading assets and liabilities to be readily incorporated in the income statement. However, the fair value hierarchy required by IAS 39 force managers to use market estimation (Level 1) to evaluate assets even when the market is illiquid 4. Thus, fair value estimates might just reflect liquidity rather than the fundamental value of these assets. Therefore, fair value accounting may introduce debt-irrelevant earnings volatility. Additionally, managers have the potential to exploit the subjectivity of fair value estimates based on proprietary models (Level 3). Such lack of verifiability alters the usefulness of accounting information for debtholders. Second, even if goodwill impairment test (IAS 36 and IAS 38) is surely a source of conditionally conservative accounting there are two caveats that work against its usefulness in debt contracting. First, IFRS 3 through the purchase price allocation provides managers with the discretion that affects the value at which goodwill is first recorded. Second, impairment test requires to periodically comparing the book value of the goodwill with its fair value. However, as stated above the estimation of fair value is highly discretionary and management can use the lack of verifiability to strategically delay a loss 5. Finally, fair values changes might originate from shocks to both cash flows and expected returns. Fair values changes that result from temporary shocks that reverse before debt maturity make earnings more volatile and thus a less efficient contracting variable. Moreover, to the extent that shock can reverse before debt maturity they are irrelevant for debtholders, because it is difficult for them to distinguish between transient and permanent shocks. Thus making fair value estimates useless for debt-contracting. 4 The same requirement applies to Investment Properties (IAS 40). 5 For example a European IT listed company on 2012 annual report shows equity which is worth around 66% of goodwill. The amortization of goodwill would have forced shareholder to recapitalize the company providing additional source of finance. Under this circumstance debtholder would be better off. 7

8 However, only equity instruments has to be necessarily evaluated at the fair value, while all other assets can be measured at the historical cost or at the fair value. Moreover, IFRS substitute in many cases unconditional conservatism with conditional conservatism, thus making firms choices particularly important. Given that shift from unconditional to conditional conservatism and the possibility to adopt or not the fair value for large categories of assets the level of conservatism, the new equilibrium, in the post IFRS adoption period would likely depend from firms characteristics and, we claim, from the presence of bankers. 2.3 Conservatism and Bankers on Board According to the agency theory corporate governance mechanisms and the board of directors perform a monitoring role, ensuring that the assets of a firm are used efficiently to guarantee the suppliers of finance a return on their investments (Shleifer and Vishny, 1997). Consistent with this view Garcia Lara et al. (2009) argue and find that strong evidence that corporate governance mechanisms increase the demand for conservative accounting numbers 6. Beekes et al. s (2004) results indicate that UK firms with a higher proportion of outside independent directors recognize bad news in a timely manner. In the same vein Ahmed and Duellman (2007) show for a sample of US firms that the percentage of inside director is negatively related to conservatism. A related stream of literature focus on board members background and shows that financial and accounting expertise improves the monitoring effectiveness (Klein, 2002; Krishnan and Visvanathan, 2008; Song et al., 2010). In this vein, Erckens et al. (2012) investigate how affiliated bankers (i.e. lender) on board influence the use of conservative accounting and document a reduction in conservatism. Their results suggest that affiliated bankers and the level of conservatism are two mechanisms that substitute each other in protecting debtholders. 6 Internal governance mechanisms such as independent board of directors have been shown to constrain aggressive practices and limiting the incidence of income increasing earnings management (Klein, 2002, Paesnell et al 2005). 8

9 While board has been mainly analyzed considering its monitoring ability and the consequences on earnings quality, less attention has been paid on the importance of boards connections and their influence on boards behavior. In this regards network theory has extensively documented the importance of information transfer trough board interlocks. Pefeffer and Salancik (1978) suggest that boards acting as boundary spanners help firms to incorporate critical information arising from the external environment into the decision making process. Haunschild and Beckman (1998) suggest that interlocks are likely to be important communication structure that facilitates information transfer. Shi et al. (2013) and Chiu et al. (2013) document that board interlocks leads to similarity of firms accounting choices. Specifically, Chiu et al. (2013) examine 118 with earnings restatement to identify extreme cases of earnings management and find that board interlocks are associated with the transmission of this behavior. Shi et al. (2013) expand this finding showing that even in non-extreme cases board interlocks leads to interlocked firms to share earnings management practices. Building on this literature we consider bankers as a source of debt-market related information. We argue that bankers possess information about the conservatism demand arising from the debt market, which becomes an informational input into the accounting decision of the firm where they serve on boards. Moreover, it has been documented that information obtained from an interlocking is likely to be highly credible (Nissbet and Ross, 1980) and trustworthy (Zajac, 1988). To be a trustworthy a source of information should be free from conflict of interests (Haunschild and Beckman, 1998). This is the reason why in our analysis we only consider interlock through non-affiliated bankers. The fact that bankers have first-hand debt-market information (Booth and Deli, 1999; Byrd and Mizruchi, 2005) and specific debt-market expertise may make other board members more comfortable with accepting a costly accounting choice, 9

10 such as conservatism. The IFRS strong tilt towards fair value accounting might have, indeed, exacerbated the conflict between debtholders and shareholders thus giving rise to an increase in the demand of conservatism (Ball et al. 2012; Khan and Watts, 2009). Therefore, bankers might turn to be particularly valuable around the IFRS information shock, especially because IFRS limit the unconditional conservatism in favor of conditional conservatism and provide discretionality in the use of fair value, making firms specific choices particularly important. In this context, bankers having debt market information are likely to be more proficient at efficient contracting and understand the benefits of conservatism. Therefore we state the following hypothesis: H1: Firms with bankers on board show more conservative accounting than firms without bankers after the IFRS information shock. 3. Data and Research Design 3.1 Sample Selection We obtain accounting and market data from Compustat Global, and analyst forecast data from the I/B/E/S international (split unadjusted) database. We begin by identifying from Compustat all public companies domiciled in Europe from 2002 to We eliminate firms in banking and financial industry (SIC codes between ) and we require at least 25 observations in each year and country to estimate conditional conservatism metric. Next, we combine accounting and market data with the analyst forecast data from I/B/E/S. To be included in the sample, we require each firm-year observation to have data available for at least one period before and one period after the mandatory adoption deadline (i.e. fiscal years beginning on or after the January 1, 2005). 10

11 Finally, we require that each firm-year observation have data necessary to calculate variables used in the regression analysis. Next, we identify mandatory IFRS adopters by retrieving information on a firms accounting standards followed from Compustat Global, and we define mandatory adopters those firms that do not adopt IFRS until it becomes mandatory 7 (i.e. fiscal-years beginning on or after 01/01/2005). The final treatment sample comprises 3,887 mandatory IFRS adopter firm-year observations from 14 European countries from 2002 to Then, we draw a control sample of GAAP firms from countries that do not require IFRS reporting during the test period selected through a propensity score matching procedure to account for differences in firms characteristics between the treated and un-treated firms while estimating the IFRS treatment effect. This yields a final control sample of 2,280 firm-year observations from 11 non-ifrs countries from 2002 to Finally, for each firm-year in the treatment sample, we manually retrieve information about the composition of the board of directors from the annual reports, and extract information about each director role, independence status, and work experiences to classify them as bankers. Information about directors primary occupation in these filings is often missing or incomplete. Hence, we collect additional information from other sources (i.e. BoardEx, Thomson One, LexisNexis, Google.com). 3.2 Conditional Conservatism Metric We operationalize conditional conservatism using the C_SCORE firm-year measure developed by Khan and Watts (2009). The C_SCORE is based on the Basu s (1997) model of asymmetric 7 A firm is classified as mandatory IFRS adopters if the data item astd in Compustat global does not equal DI prior to fiscal year beginning on or after January 1,

12 timeliness to estimate a firm-year measure of accounting conservatism. Following Jayaraman (2012) we estimate the Khan and Watts s model for each country and year requiring at least 25 observations for each group. The Basu model is specified as: (1) where EARitk is the net income before extraordinary items scaled by lagged market value of equity of firm i in year t in country k, Ditk is a binary variable that equals one if Ritk is negative and zero otherwise, Ritk denotes annual returns obtained by cumulating monthly returns starting from the fourth month after the firm s fiscal year end. The coefficient on Ritk (β3) measures the timeliness of earnings with respect to good news (positive returns); while the coefficient on the interaction between Ritk*Ditk (β4) captures the incremental timeliness of earnings with respect to bad news (negative returns), or conditional conservatism. To estimate the timeliness with which accounting reflects good and bad news at the firm-year level, Khan and Watts (2009) specify that both the timeliness of good news (referred to as the G_SCORE) and the incremental timeliness of bad news (C_SCORE) are linear functions of firmspecific characteristics draws from factors that explain cross-sectional variation in the demand for accounting conservatism: (2) (3) 12

13 C_SCORE is the firm-year measure of conditional conservatism, while G_SCORE is the firmyear measure of good news timeliness. Equations (2)-(3) are not regressions models but equations. We thus substitute them into equation (1) to obtain equation (4), after augmenting the model with the additional interaction terms between returns and firms characteristics 8 : (4) C_SCORE and G_SCORE vary across firms as well as over time within a country through crosssectional variation in the firm-year characteristics (i.e., size, market-to-book and leverage). 3.3 Research Design We examine how the presence of bankers is related to the extent of accounting conservatism around IFRS mandatory adoption by employing a panel dataset with firm-year level observations. Our research design encompasses three steps. First, we use a treatment sample of mandatory IFRS adopters firm-year observations and a control sample of non-ifrs adopters firm-year observations selected through a propensity score matching procedure. Specifically, our final sample consists of firms from the EU countries that adopt IFRS reporting from 2005, and from countries that do not mandate IFRS reporting over the entire test period identified through a propensity score-matching sample. The use of this benchmark allows us to account for global time-trends that could affect accounting conservatism, while the matching procedure minimizes the degree of heterogeneity between the treated and un-treated firms. To build up the propensity 8 We delete observations in the top and bottom one percent of EAR, R, SIZE, M/B, and LEV. 13

14 score matching sample we first take the firm-specific mean using data between 2002 and 2004, which is before the IFRS mandatory adoption deadline, of the following variables: SIZE, LEV, ROA, GROWTH, FOLLOWING. Then, we estimate the propensity score by running a logistic regression of the binary variable for being a firm from an IFRS-adoption country on the firmspecific mean of those variables as well as industry (Campbell 1996) and country fixed effects, using one observation for each firm 9. Next we match, without replacement, each IFRS firm with a non-ifrs firm, which has the closet predicted value from the propensity score matching regression. The use of a propensity score matching technique, by defining the degree of similarity across treated and untreated firms alleviates concerns related to the assumption of parallel patterns in the counterfactual scenario of no information shock in a difference-in-differences design. In this way, we can capture the change in accounting conservatism in the pre- versus post-mandatory adoption period for mandatory adopters relative to the change for the benchmark firms, taking into account unobserved heterogeneity across firms or time-invariant selection bias. This issue is particularly severe in corporate governance research, to the extent that the presence of a banker on board is endogenous. In our setting, there may be many reasons for board composition and financial reporting quality to be jointly determined by some unobserved firm characteristics. If these unobserved firm characteristics are time invariant, then this design which exploits IFRS mandatory adoption as an exogenous shock addresses simultaneous determination problems. 9 Propensity-score matching models (Rosenbaum and Rubin 1983) match observations with respect to the probability of be treated, which in our setting is the likelihood of being a firm from an IFRS country Propensity score matching models seem to be particular suitable in our setting. First, this approach creates samples in which IFRS and not adopters are similar, providing a good framework to assess how IFRS affect conservatism. Second, selection or treatment effect models (Heckman 1979) rely on a specific functional form. Matching models do not rely on a specific functional form. However, this approach does not take into account unobservable heterogeneity across firms in estimating treatment effects. 14

15 Second, we identify firms that have a banker on board. To do this, we code up a binary variable (BANKER) marking firms with a banker on board both in the pre and in the post IFRS mandatory adoption period (i.e and 2006 fiscal years). In this way, we employ a timeinvariant indicator that groups firms for which the banker representation is stable over time. Untabulated results indicate that only few firms that have a banker on board in the pre-ifrs mandatory adoption period do not have it more in the post-ifrs. The combination of the stickiness of banker representation and a differences-in-differences design allow us account for the endogeneity in board composition. The last element of our research design includes an extensive fixed-effects structure. We include country, industry and year fixed effects to take into account the effects of potentially confounding events around IFRS mandatory adoption. In this way we control for general trends or common shocks unrelated to IFRS. We thus propose the following estimation model (without firms and time subscripts): C_SCORE= β0 + β1ifrs + Ʃj βj CONTROLS + Ʃi βi FIXED EFFECTS+ γ (5) where C_SCORE stands for the conditional conservatism metric developed by Khan and Watts (2009). IFRS is a binary variable that takes the value of one for firms that apply IFRS only when it becomes mandatory in 2005 for fiscal-years beginning on or after 01/01/2005, zero otherwise. CONTROLS denotes a broad set of control variables (see Section 3.4), while FIXED EFFECTS represents country-industry and year fixed effects or in alternative specifications year and firm fixed effects. 15

16 In equation (5) we do not include indicator variables for the presence of a banker on board. Rather, in the empirical analysis we add to equation (5) a set of non-overlapping binary variables to estimate separate IFRS effects on accounting conditional conservatism considering the presence of a banker on board. For instance, when we estimate differential changes in accounting conservatism around IFRS adoption for firms with a banker on board and without a banker on board we propose the following model (without firm and time subscripts): C_SCORE= β 0+ β 1IFRS*Bankers + β 2 IFRS*Non-Bankers +Ʃj βj CONTROLS +Ʃi βi FE + Ɛ (6) In this model we replace the single IFRS indicator from equation (5) with two non-overlapping binary indicator, one for firms with a banker (IFRS*Bankers) and one for firms without a banker (IFRS*Non-Banker). In this way, we can compare the IFRS effects on accounting conservatism across the different groups of treatment firms to disentangle the role of bankers on board on accounting conservatism around IFRS mandatory adoption. 3.4 Control Variables The models include year-industry (using the Campbell (1996) industry classification) and country fixed effects. Our main inference is based on heteroskedasticity-corrected standard errors, which are adjusted at country-level clustering (Daske et al. 2008) 10. In addition, we include a set of controls variables that capture cross-sectional variation in the demand for accounting conservatism (Watts 2003). 10 We also report standard errors adjusted at country level, which allows for correlation across time but is more conservative than firmlevel clustering (Daske et al., 2008). In untabulated analysis we used firm- level clustering (Gow et al 2010; Bell and Mc Caffrey, 2002) and the results are substantially unchanged. 16

17 We control for loss-reporting firms by coding up a binary variable (LOSS) equals to one whether the I/B/E/S earnings per share is less than zero, zero otherwise. We take into account firm performance using return on assets (ROA), computed as net income over the end of the year total assets. We control for the firm size (SIZE) using the log of the firm total assets at the beginning of the year, as larger firms have a lower contracting demand for accounting conservatism. We include growth prospectus using the market to book ratio (MB) to the extent that growth prospectus is found to increase the demand for conservatism. LEVERAGE, which is the sum of long-term debt and the current portion of long-term debt scaled by total assets, takes into account debt-contracting demand for conservatism, in the same vein we control for the percentage change in total liabilities (DISSUE). We account for external monitoring mechanisms that may be related both with accounting conservatism and un-affiliated banker representation. FOLLOWING is the residuals from a regression of the number of analysts following a firm in a given year on firm size. SD_IB captures earnings volatility and is computed as the standard deviation of earnings before interest and taxes. We control for fundamental uncertainty using: the investment cycle (INV_LENGHT) computes as depreciation expenses deflated by total assets; the degree of intangibility (INTANGIBILITY) calculated as one minus the value of plant, property, and equipment as a fraction of total assets to the extent that intangible assets are more difficult to evaluate (Harris and Raviv, 1991) thus increasing the demand for accounting conservatism; and the annual standard deviation of the monthly stock returns (SD_VAR). 4. Empirical Results 4.1 Descriptive Statistics 17

18 Table 1 illustrates the sample distribution by country. The number of observations varies widely across countries: Austria has the lowest number of observations (5) and the UK has the highest (1,593). France is the country with the highest bank representation (32.4 percent), while neither Austrian nor Danish firms have bankers on board. Table 2 presents descriptive statistics of the variables used in the regression analyses. Panel A report descriptive statistics for the overall sample. The mean (median) value of C_SCORE is (-0.002); this value seems to be consistent with Jayaraman (2012). We also perform additional tests to verify to what extent the Khan and Watts measure (C_SCORE) is appropriate in our setting (see Appendix A). We next split the overall sample into two groups: IFRS adopters (Panel B) and Non-IFRS adopters (Panel C). The propensity score model appears effective in forming a balance sample of treated and non-treated firms. The control variables have a sample distribution consistent with prior literature. [INSERT TABLES 1 AND 2 ABOUT HERE] 4.2 Main analysis on conditional conservatism We start our empirical analysis by exploring the average effect of the information shock on conditional conservatism. We use cross-sectional, time-series panel regressions, which benchmark firms in IFRS adopting countries against firms in non-ifrs countries. Table 3, model (1) reports results from OLS regressions with robust standard errors that are clustered by firm. In model (2), Table 3, we estimate the average effect model replacing the country and industry fixed effect with firm fixed 18

19 effect. In this way we can control for time-invariant and potentially unobserved differences between firms. We find that the on average the information shock, given by the mandatory IFRS adoption, has an effect on the level of accounting conservatism, when we use the firm-fixed effects structure In models (3) and (4) we investigate whether there is heterogeneity across firms with respect to the change in conservatism around the IFRS adoption. We replace the IFRS indicator with two non-overlapping indicator variables, one for firms with a banker on board (i.e. IFRS*Banker), and one for firms without bankers (i.e. IFRS*Non-Banker). The results are presented in columns (3)-(4). The estimated coefficient on IFRS*Banker is positive and significant (0.076, p<0.000), while the estimated coefficient on IFRS*Non-Banker is not significant at any conventional level (0.005, p>0.1). These results hold even when we use a more conservative structure (i.e., firm fixed effects, model (4)). Overall, evidence from models (3) and (4) supports our main hypothesis. A significant increase in the level of conservatism for firms with banker representation follows the mandatory IFRS adoption. Consistent with the above evidence contrast F-test indicate that the coefficient for IFRS*Banker and IFRS*Non-Banker are significantly different. [INSERT TABLE 3 ABOUT HERE] 4.3 Analysis of country-level incentives Debt Market Importance A potential concern for this study is that the presence of bankers might just capture the importance of the debt market in a given country. To rule out this alternative explanation we 19

20 estimate a model that takes into account the relevance of the debt market for the demand of accounting conservatism. To proxy for the importance of the debt market we use a conditional variable (CV) that is the sum between private and public bond market capitalization over grossdomestic product for the year 2004, taken from the World Bank. We next transform this measures (DEBT_MARKET) into a binary variable (HIGH_DEBT_MARKET) based on whether a country specific value is above or below the treatment sample country median. Next, we create four non-overlapping variables: IFRS*Banker*HighCV is a dummy variable equals to one for firms reporting under IFRS with bankers on board (i.e., BANKER equals to one) with the conditional variable above the sample median(i.e. HIGH_DEBT_MARKET equals one); IFRS*Non-Banker*LowCV is a dummy variable equals to one for firms reporting under IFRS without bankers on board (i.e., BANKER equals to zero) with the conditional variable below the sample median (i.e. HIGH_DEBT_MARKET equals zero); IFRS*Banker*LowCV is a dummy variable equals to one for firms reporting under IFRS with bankers on board (i.e., BANKER equals to one) with the conditional variable below the sample median (i.e. HIGH_DEBT_MARKET equals zero); IFRS*Non-Banker*HighCV is a dummy variable equals to one for firms reporting under IFRS without bankers board (i.e., BANKER equals to zero) with the conditional variable above the sample median (i.e. HIGH_DEBT_MARKET equals one). Table 4, column (1)-(2) presents the estimated coefficients and t-statistics. We find a significant increase in the level of conservatism for firm with banker on board domiciled in countries where the debt market is likely to be more important. Turning on countries where debt market is less important, we document a significant increase in accounting conservatism only for firms with a banker on board. However, it is important to notice that the magnitude of the coefficient in the two groups (IFRS*Banker*HighCV and IFRS*Banker*LowCV). When a banker is on board the 20

21 coefficient is almost double. This result implies that once IFRS became mandated, only firms with bankers on board experience an increase in the level of accounting conservatism, irrespectively from the importance of the debt market Importance of the IFRS shock to Conservatism The main idea of our study is that IFRS alter financial reporting in ways we conjecture reduce their usefulness in debt contracting because of the wider use of fair value. If IFRS adoption caused a change in financial reporting, this change should be a function of how much a country s accounting standards are altered by mandatory IFRS adoption. To test this possibility, we classify countries according to the extent to which IFRS result in less conservative financial reporting relative to local accounting. This allows us to take into account differences in the magnitude of the information environment shock across countries driven by the extent to which recognition and disclosure rules change (Ball et al. 2012). To proxy for the differences between IFRS and local GAAP in term of accounting item we use the conditional variable (CV) computed as the sum of six items (ACC_DIFF). We identify six important accounting standards that might have changed the level of conservatism (see Appendix 1 for an explanation). Appendix A1 provides both the list of the six accounting standards used in measuring the variable (ACC_DIFF) and the individual country scores for each item. The table indicates which countries have been coded 1 (-1), indicating that the mandatory IFRS adoption mechanically lead to an increase (decrease) in the level of conservatism. If the country it is coded 0 it means that the local GAAP required the same accounting treatment as the IFRS. A score of 0.5 (-0.5) indicates that the local GAAP provides the firms with the possibility of an accounting 21

22 treatment similar to the IAS, thus the impact/ magnitude of mandatory IFRS adoption is not, in average, as strong. We then partition the treatment sample using the treatment sample country median. Countries above the median are classified as countries with greater differences (HIGH_ACC_DIFF equals to one). Next, we create four non-overlapping variables: IFRS*Banker*HighCV is a dummy variable equals to one for firms reporting under IFRS with bankers on board (i.e., BANKER equals to one) with the conditional variable above the sample median(i.e. HIGH_ACC_DIFF equals to one ); IFRS*Non-Banker*LowCV is a dummy variable equals to one for firms reporting under IFRS without bankers on board (i.e., BANKER equals to zero) with the conditional variable below the sample median (i.e. HIGH_ACC_DIFF equals zero ); IFRS*Banker*LowCV is a dummy variable equals to one for firms reporting under IFRS with bankers on board (i.e., BANKER equals to one) with the conditional variable below the sample median (i.e. HIGH_ACC_DIFF equals zero); IFRS*Non-Banker*HighCV is a dummy variable equals to one for firms reporting under IFRS without bankers board (i.e., BANKER equals to zero) with the conditional variable above the sample median (i.e. HIGH_ACC_DIFF equals one). Table 5 columns (3) and (4) present the results. We find that bankers have a significant a positive effect on the level of conservatism once the IFRS became mandated. [INSERT TABLE 4 ABOUT HERE] 4.3 Analysis of firm-level incentives So far, our evidence supports our explanation and suggests that firms with banker representation experience an increase in the level of accounting conservatism after the mandatory IFRS 22

23 adoption. However, we need to examine whether the increase in the level of accounting conservatism is driven by firm-specific characteristics rather than the banker representation. To explore this point we replicate our main analyses by augmenting equation (6) with firm-level conditional variables that capture: i) firm-level incentive to increase the accounting conservatism and ii) the intensity of the conflict between share-and bond-holders over dividend policy. To proxy for firm-level incentive to increase the level of conservatism, we use the change in the amount bond to the extent that bondholders are more likely to rely on public information than private debt-market participant. Following Ahmed et al. (2002) we focus on change in leverage and change in total dividend as proxies for conflicts over dividend policy 11. Our last partitioning variable is a proxy for the level of information asymmetry between firms and providers of finance based on the quality of analyst information environment in the years (Duchin et al., 2010). Our first firm-level partitioning variable is calculated as the change in the total amount of bond issued between the pre- and the post-ifrs mandatory adoption date. Data on the bond issued are drawn from DEALSCAN. To the extent that the 2005 is the switching year, we first calculate the mean of the total amount of bond issued for the years and for the years Then, we subtract for each firm the average for the pre-ifrs period from the average for the post-ifrs period. Finally, we create a binary variable (ΔBond) based on the sample distributions of the changes around IFRS mandatory adoption. Specifically, ΔBond takes the value of one if a firm experience an above the median change in the total amount of bonds between the pre- and post IFRS mandatory adoption. 11 As Ahmed et al. (2002) propose we also use the standard deviation of return on assets. Results (un-tabulated for sake of brevity) are unchanged with those reported. 23

24 Our second firm-level partitioning variable is calculated as a rolling average over the previous three years (i.e., years t, t-1, t-2) of the ratio between total liabilities and total assets (i.e. LEV). Next, we partition the treatment sample focusing on the changes in LEV around IFRS adoption. To do so, we subtract for each firm the rolling average in year 2004 from the rolling average in year Next, we create a binary variable (ΔLeverage) based on the sample distributions of the changes around IFRS mandatory adoption. Firms with above sample median value of ΔLEV are classified as firms having strong bondholders-shareholders conflicts. We calculate our third reporting incentive variable (i.e ΔDividend) as a rolling average over the previous three years (i.e., years t, t-1, t-2) of the ratio between total dividend and total assets. Next, we partition the sample focusing on the changes in DIV around IFRS adoption. To do so, we subtract for each firm the rolling average in year 2004 from the rolling average in year Next, we create a binary variable (ΔDIV) based on the sample distributions of the changes around IFRS mandatory adoption. Firms with above sample median value of ΔDIV are classified as firms having strong bondholders-shareholders conflicts. The last firm-level partitioning variable (i.e., Information Cost) is based on the degree of information asymmetry between firms and providers of finance and exploits analyst earnings forecast data (Duchin et al. 2010). We consider the availability, accuracy, and homogeneity of analyst earnings forecasts. The first is the number of analysts who release at least one earnings forecast during the year. The second is the forecast error computed as the absolute value of the analyst forecast accuracy, deflated by the stock price at the beginning of the fiscal year: ACCURACYit = Actual Earningsit Median Forecastit /Stock Priceit,, where Actual Earningsit is the Actual I/B/E/S annual EPS for firm i in year t, Median Forecastit is the median of forecasts made by analysts in our sample from the 11th month of the fiscal year to 3 days before the annual earnings announcement for firm i and year t, and Stock Priceit is the stock 24

25 price of firm i at the end of year t. 12 Using the same forecasting window (from the 11th month of the fiscal year to calculate to 3 days before the annual earnings announcement) we compute the homogeneity of analyst forecasts as the Standard Deviation of Forecasts/Stock Price. Then we compute the mean over of each of those proxies and we combine them in one indicator by averaging a firm s percentile ranking in the sample according to each measure (for the number of analysts, the reverse ranking is used). Firms with above the sample median of the index are classified as having higher level of information asymmetry between firms and provider of finance (Information Cost equals to one). Table 5 reports the estimations results. Overall, we find an increase in the level of conservatism for those firms with an above the median change in the conditional variable and with a banker on board. When we use the firm fixed effect structure (Model (2) (4) (6) and (8)) we find evidence that firm-level characteristic matter in leading to an increase in conservatism. However, it is worth notice that the magnitude of the estimated coefficient on IFRS*Banker*HighCV is double than the estimated coefficient on IFRS*Non-Banker*HighCV. [INSERT TABLE 5 ABOUT HERE] 5. Additional Analyses In this section, we conduct a series of additional analyses to check the robustness of the above results. 5.1 Alternate control sample 12 We remove the effect of stale forecasts by employing only the last forecast each analyst made in this window if they issue more than one forecast. 25

26 The above analyses include as control sample firms from countries that do not allow IFRS reporting. In untabulated results we use voluntary IFRS adopter as an alternate control sample. Results are still qualitatively the same. 5.2 Placebo analysis To reduce the concern that our results capture general effects unrelated to the information shock we conduct a placebo analysis. (1) We replicate our analyses by counterfactually varying the event year of IFRS mandatory adoption. By this placebo analysis we intend to appraise the timing of the treatment IFRS event. To do this, we shift the date of the IFRS mandatory adoption (t = 0) to different dates, from year t-2 to year t+2. Then, we replicate our main analyses. If mandatory IFRS reporting significantly concurs to shape accounting conservatism, we should not observe significant changes C_SCORE metric before or after This is what we document in untabulated results 5.3 Event study in the post-ifrs mandatory adoption period Our results provide a quite consistent picture about the role of banker in shaping the level of accounting conservatism. We exploit IFRS adoption as an exogenous shock that may trigger a debtholder- shareholder conflict and find that firms with a banker are more prone to increase the level of accounting conservatism after IFRS adoption. The crucial assumption of the paper is that IFRS adoption effectively triggers a potential conflict between debtholder and shareholder. To assess the robustness of our results we carry out an event study of conservatism change around a significant increase in idiosyncratic uncertainty (Khan and Watts 2009). Khan and Watts (2009) argue that a high level of uncertainty by increasing the difficulty in forecasting the magnitudes 26

27 and timing of future cash flows should generate a higher demand for conservatism. Therefore, we are interested in examining whether firms with bank representation react to a firm-specific shock in the demand for accounting conservatism differently from firms without a banker on board. In this test we only consider the period and we measure the idiosyncratic uncertainty as the annual standard deviation in monthly stock returns. Specifically, we take the change from a year to another of our idiosyncratic uncertainty proxy, and we define a significant increase as one above the 90th percentile of the distribution of the annual changes in the idiosyncratic uncertainty proxy in the years Then, we regress the annual change in accounting conservatism on indicators variables marking the year before the shock (i.e. t-1), the event year (i.e. t), and the year subsequent the shock (t+1). We run separately this regression for firms without a banker on a board and for firms with a banker on board using as control sample firms without a banker on board in the 10th percentile of the distribution of the annual changes in the idiosyncratic uncertainty in the years Table 6 presents the estimation results, while figure 1 depicts the graph of the event study. The vertical axis presents the mean of the change in the conservatism metric, while the horizontal axis shows the event year. Figure 1 shows that accounting conservatism increases around a significant increase in information uncertainty consistently with the findings in Khan and Watts (2009). More interestingly, firms with a banker on a board react to this shock with a larger increase in accounting conservatism than firms without a banker on a board. This evidence implies that a sophisticated financial expert as a banker leads firms to timely adjust the supply of conservatism following an informational shock. [INSERT TABLE 7 ABOUT HERE] [INSERT FIGURE 1 ABOUT HERE] 27