We examine whether unconditional conservatism supported stock valuations during the COVID-19 crisis. Building on evidence that this form of conservatism improves downside-risk assessment and mitigates bankruptcy and litigation concerns, we hypothesize that its benefits were amplified during this period of heightened uncertainty, sustaining valuations and reducing investor uncertainty.
We analyze stock price patterns of US-listed firms around the onset of COVID-19. Following prior literature, we employ a composite measure based on three widely used proxies to measure each firm’s level of unconditional conservatism in the preceding three fiscal years (2017–2019).
Consistent with our hypothesis, higher conservatism resulted in superior abnormal stock returns during the crisis. These benefits were especially pronounced in industries more exposed to COVID-19 and among firms with greater reliance on international sales, which made them more susceptible to global disruptions. In contrast, the advantages of conservatism were less evident in firms with stronger corporate governance, greater financial stability, more transparent reporting, and steadier operating performance. We also find that conservatism resulted in lower volatility and narrower spreads during the crisis and was linked to fewer negative special items in 2020, suggesting reduced vulnerability to write-down losses stemming from the pandemic.
Our findings suggest that unconditional conservatism, often seen as detrimental to valuation, is beneficial amid significant market uncertainty. We contribute to the long-standing debate on the desirability of conservatism in financial reporting, underscoring the critical role that accounting practices can play for investors and the broader economy.
1. Introduction
This paper examines whether unconditional conservatism, often viewed as distorting valuation, provided informational benefits to investors during the COVID-19 crisis, thereby supporting stock market performance amid heightened uncertainty. Unconditional conservatism refers to accounting practices that systematically favor a cautious valuation of net assets due to uncertainty about their future benefits. These practices include accelerated depreciation, immediate expensing of R&D or asset improvement costs, or last-in-first-out inventory valuation. Unlike conditional conservatism, which triggers asset write-downs in response to adverse news, unconditional conservatism is applied at an asset’s inception, regardless of future developments. This results in a consistent understatement of net assets and earnings, creating a protection against future uncertainties and potential losses. Consequently, when adverse news occurs, it may already be partially reflected in financial statements.
Conservatism is a long-standing and pervasive characteristic of accounting. Empirical evidence shows that conservative accounting practices exist across various countries and institutional settings (Ball et al., 2000, 2003; Bushman and Piotroski, 2006; Pope and Walker, 1999). Examples of such practices can be found throughout history, even dating back to the late medieval period (Basu, 2009; Chatfield and Vangermeersch, 1996; Littleton, 1941; Penndorf, 1930). Given its deep roots, the debate over the desirability of conservatism in financial reporting has persisted for decades, leading scholars to refer to it as the “eternal debate” (Orthaus et al., 2023).
A prominent view in this debate suggests that conservatism endures because it benefits certain stakeholders in specific contexts. For example, it can shield executives and auditors from litigation exposure, reduce a firm’s tax burden, protect corporate resources from expropriation by corrupt officials, or obscure excessive profitability in industries vulnerable to government intervention (Cano-Rodríguez, 2010; Chang et al., 2025; Mensah et al., 1994; Qiang, 2007). Moreover, conservatism can reassure lenders about the risk of excessive distributions to shareholders, preserving creditors’ interests (Ahmed et al., 2002; Watts, 2003).
However, in recent decades, both regulators and segments of the academic community have increasingly opposed it. Critics argue that conservatism biases the information provided to stock market investors, viewed as among the primary users of financial statements, who would benefit more from neutral financial reporting. In response to this stance, a body of empirical research has emphasized the informational and contractual advantages of conditional conservatism. Nevertheless, criticism of unconditional conservatism persists even within this literature. For instance, Ball and Shivakumar (2005) argued that while conditional conservatism improves contracting by ensuring timelier recognition of bad news, introducing an unconditional bias in financial statement numbers can undermine decision-making and reduce efficiency. Consequently, most studies on the benefits of conservatism for the stock market have focused exclusively on the conditional form, often suggesting that unconditional conservatism is either detrimental or, at best, irrelevant to shareholders’ information needs. Given this perspective, the potential for unconditional conservatism to positively influence stock market information has remained largely unexplored.
In this paper, we address this gap by examining the influence of unconditional conservatism (hereafter just “conservatism,” unless differently specified) on stock market behavior during the early months of 2020, when the COVID-19 crisis broke out. While conservatism’s impact on earnings predictability may hinder price discovery in ordinary times, we propose that during periods of spiking uncertainty triggered by adverse systemic events, it can facilitate price discovery as it provides market participants with cautious valuations of a company’s uncertain net assets, which are particularly relevant when downside risks are likely to materialize, and insolvency concerns arise.
As a period of significant global uncertainty that had profound effects on the stock market (Bhattacharya, 2024), the onset of the COVID-19 pandemic provides an ideal context to examine the informational benefits of conservatism, which is often regarded as accounting’s response to uncertainty in business operations (Kale and Villupuram, 2025; Penman, 2021). As the economic implications of the pandemic became clearer and expectations shifted toward a more prolonged and uncertain recovery, major U.S. stock indices experienced sharp declines, with some falling over 12% in a single day (Pisani, 2021). During this period of heightened market volatility and rapidly evolving information, investor uncertainty regarding the potential extent of the downturn further contributed to market instability (Imbert, 2020; Frazier, 2021). In the following weeks, the markets unexpectedly began to regain ground as government and monetary institutions worldwide introduced significant stimulus measures to stabilize the economy. As is typical in the aftermath of market crises, the path to recovery varied across firms, depending on investors’ judgments of which companies were better positioned to withstand the downturn (Mahata et al., 2021).
We hypothesize that conservatism influenced these judgments by reducing investor uncertainty about the pandemic’s impact on firms’ cash flow prospects and their ability to manage the crisis, thus improving cumulative returns, lowering return volatility, and reducing trading risks. We identify two channels through which conservatism achieved these effects.
First, the rapid onset of the COVID-19 pandemic disrupted business plans across industries and market segments, causing cash flow expectations to fall rapidly. Investors were faced with the possibility of near-term write-downs of uncertain scale, heightening awareness of the downside risks associated with corporate investments (e.g. the potential for earnings to decline significantly if expected returns and future revenues did not materialize).
In this context, conservative financial reporting may have played a material role in reducing investor uncertainty. By systematically understating uncertain net assets, conservative accounting practices mitigate the risk of overestimating a firm’s financial health. This likely helped investors better assess downside risk by providing a more cautious view of corporate financial conditions for stock valuation. This argument aligns with Penman and Zhang (2021), who suggested that conservatism enhances the informativeness of the book rate of return regarding the risk inherent in corporate investments. Similarly, Kale and Villupuram (2025) argued that conservative financial information aids in evaluating equity market risk, particularly downside risk, during periods of heightened uncertainty. Earlier research also supports the idea that conservatism aids a more realistic assessment of downside risk in uncertain net assets. For example, Mashruwala and Mashruwala (2018) found that the stock valuations of less conservative firms are disproportionately influenced by overly optimistic investor beliefs about future payoffs. Likewise, Hirshleifer et al. (2004) observed that recognizing uncertain net assets on the balance sheet can lead to investor mispricing, especially among those with limited attention.
Second, the swift decline in cash flow prospects and rising corporate debt levels significantly heightened insolvency risk (Friesenhahn and Kwan, 2020), making investors and analysts look at firms’ solvency and cash-generating ability with renewed attention as the pandemic unfolded (Kent et al., 2023). These growing concerns heightened the probability of financial distress within the business sector, prompting questions about the capacity of the U.S. court system to manage a possible rise in related proceedings (Greenwood et al., 2020). Conservatism can alleviate investors’ concerns about bankruptcy risk by ensuring that corporate resources are preserved within the company, preventing incautious distributions, and mitigating managerial opportunism in financial reporting (Biddle et al., 2022). Additionally, conservatism can reduce the uncertainty related to litigation exposure (Qiang, 2007), a risk that becomes particularly acute in volatile markets or during sudden price drops (Kim and Skinner, 2012), which can further exacerbate investors’ concerns about insolvency (Arena, 2018).
To test our hypothesis, we analyze the stock price patterns of US-listed firms at the onset of the COVID-19 crisis, measuring their conservatism levels during the three preceding fiscal years (2017–2019). Building on prior conservatism literature (Aier et al., 2014; Chen et al., 2014; D’Augusta et al., 2016; Ettredge et al., 2016; Hui et al., 2009; Kim et al., 2013; Mashruwala and Mashruwala, 2018), we use a composite proxy that incorporates three widely used measures of conservatism: the estimated balance sheet reserves created by conservatism, the persistent understatement of equity book values relative to market values, and the average magnitude of negative non-operating accruals.
Our findings reveal that conservatism is positively associated with cumulative abnormal returns over the forty-five days following the World Health Organization’s (WHO) declaration of the pandemic on March 11, 2020. Additionally, conservatism is negatively associated with the surge in bid-ask spread and abnormal return volatility over the same 45-day period. These results suggest that conservatism mitigated investor uncertainty as the pandemic unfolded, thereby reinforcing stock performance.
Supplemental analyses indicate that the positive effect of conservatism on stock returns is persistent, with no evidence of reversal over time. In firms with low initial conservatism, its magnitude declines in later crisis stages, consistent with the depletion of protective reserves. In addition, the effect of conservatism is attenuated for domestic firms, which are less affected by global disruptions, industries less exposed to COVID-19, and firms with stronger governance (e.g. independent boards), suggesting that conservatism substitutes for other monitoring mechanisms. Furthermore, the benefits of conservatism are smaller where fundamentals are strong but they are amplified among financially distressed firms. Finally, less conservative firms recorded more negative special items in 2020, indicating greater asset write-downs and reinforcing conservatism’s role in mitigating investor concerns about near-term losses.
Our study makes several key contributions to the literature. First, we provide evidence of the positive effects of unconditional conservatism on stock market investors, challenging the prevailing view that conservatism’s benefits are confined to the conditional form. Specifically, we show that unconditional conservatism can positively influence stock market behavior by reducing investor uncertainty about firm valuations during macroeconomic crises. Second, our study enhances the understanding of the factors that drove stock price fluctuations and market dynamics during a period of significant economic and financial turbulence. Given that spiking uncertainty and sudden price drops can have far-reaching consequences for various stakeholders beyond just stock traders, our findings underscore the broader role that accounting can play in stabilizing the economy. Finally, our results contribute to the long-standing debate on the desirability of conservatism, suggesting that the regulatory tendency to view it as a negative feature of accounting may be misguided, especially when considering its benefits during periods of macroeconomic uncertainty.
The remainder of the paper is organized as follows. Section 2 discusses the theoretical background and develops the hypothesis. Section 3 describes the sample selection, variable measurement, and empirical design. Section 4 presents the main results, and Section 5 provides additional analyses. Section 6 concludes.
2. Theoretical background and hypothesis
Conservatism has long been recognized as a foundational principle of accounting. In recent decades, however, both regulators and parts of the academic community have increasingly opposed it. Accounting standards emphasize that the purpose of financial reporting is to provide investors with information for valuing firms and making investment decisions (FASB, 2010). Many scholars argue that conservatism undermines this goal by distorting shareholder information, which would be better served by neutral, bias-free reporting [1]. This view has also gained traction among regulators, who oppose conservatism for its perceived negative impact on shareholder information (FASB, 2010; Watts, 2003).
Despite the ongoing debate about conservatism’s merits, empirical examination of its effects on stock market outcomes remained limited for many years. In a literature survey published toward the end of the 2000s, Armstrong et al. (2010, p. 190) observed that conservatism’s potential to assist investors in valuation had been largely overlooked, perhaps due to “the common perception (or, possibly, misperception) that although conservative accounting may assist firms in contracting and governance settings, such benefits come at the expense of accounting’s role in valuation.”
In response to this gap, a growing body of empirical research over the past 2 decades has examined conservatism’s impact on stock markets (D’Augusta, 2024). Several studies suggest that conservatism can negatively affect valuations and hinder price discovery. Penman and Zhang (2002), for example, argued that conservatism creates hidden reserves during investment growth that reduce earnings persistence, potentially leading to mispricing if investors fail to recognize this effect. Similarly, Mensah et al. (2004) found that conservatism increases earnings volatility and forecast dispersion, while Narayanamoorthy (2006) observed that investors may overlook the time-series properties of earnings shaped by conservative accounting.
Other studies show that conservatism can further complicate investment decisions. Chen et al. (2014) argued that reduced earnings persistence may lead investors to discount conservatively reported earnings, offsetting benefits such as contracting efficiency or lower litigation risk. Kim et al. (2019) found that analysts, especially less experienced ones, often fail to adjust for conservatism, producing biased valuations. D’Augusta and Redigolo (2019) similarly noted that analysts may require more management guidance to interpret its effects. Consistent with these findings, Heflin et al. (2015) showed that conservatism in GAAP earnings prompts users to rely more heavily on non-GAAP measures, which may provide a better signal for investors.
Collectively, these studies suggest that conservatism poses challenges for stock market investors, potentially obscuring the economic performance of firms. However, a contrasting strand of research argues that the conditional form of conservatism (i.e. the timely recognition of bad news in earnings) can yield beneficial contracting effects that ultimately serve investor interests. According to Ball and Shivakumar (2005, p. 92), while “unconditional conservatism seems at best neutral and possibly inefficient,” conditional conservatism can improve contracting efficiency and reduce information asymmetry, thus benefiting investors.
Numerous studies have provided empirical support for the positive effects of conditional conservatism on stock market outcomes. Research has shown that timely loss recognition can lower information asymmetry, reduce investor uncertainty, mitigate IPO underpricing, decrease the cost of equity, and ultimately improve stock performance (García Lara et al., 2011, 2014; Kim and Pevzner, 2010; Kim and Zhang, 2016; Kim et al., 2013, 2024). Other studies have highlighted the benefits of conditional conservatism in enhancing the quality of voluntary disclosures, which can further assist investors in valuation (D’Augusta, 2022; D’Augusta and DeAngelis, 2020).
However, aligning with Ball and Shivakumar’s (2005) conjecture that informational benefits pertain exclusively to the conditional form, the potential for unconditionally conservative accounting to positively influence stock market valuations remains underexplored. This gap in the literature has left the impression that unconditional conservatism is either irrelevant or detrimental to investor valuation needs. Indeed, most studies focused on the COVID-19 crisis have primarily examined the conditional form of conservatism, similar to the general tendency of the literature analyzing the benefits of conservative accounting for the stock market. For instance, D’Augusta and Grossetti (2023) found that this form of conservatism helped investors interpret earnings announcements by U.S. firms during the pandemic. Likewise, focusing on Chinese firms listed on the Shanghai and Shenzhen Stock Exchanges, Cui et al. (2021) showed that conditionally conservative reporting was associated with smaller declines in stock performance during the outbreak. Earlier studies on conservatism’s benefits during the 2008 financial crisis, highlighting better market performance, more accessible credit, or reduced investment curtailments (Balakrishnan et al., 2016; Kim and Shawn, 2022; Zhang, 2020), have also primarily considered conditional conservatism. An exception is Francis et al. (2013), who, in a robustness check, found that proxies for both forms of conservatism are positively associated with buy-and-hold returns during 2008 and 2009.
Overall, the literature reviewed in this section underscores the need to explore how unconditional conservatism affects the stock market during crises. Our research addresses this gap by expanding the understanding of conservatism’s role during periods of heightened economic uncertainty. We argue that unconditional conservatism, despite concerns about its potential to distort valuation, may offer informational benefits in such contexts. By embedding caution into reported net assets before adverse events occur, it can reduce investors’ uncertainty about firms’ financial health and downside risk. These effects are likely to be particularly relevant during crises, when cash flow expectations decline rapidly and insolvency concerns intensify. Accordingly, we expect that firms exhibiting higher levels of unconditional conservatism prior to the outbreak experienced higher cumulative abnormal stock returns during the early phase of the COVID-19 crisis. We formalize this expectation in the following hypothesis:
Unconditional conservatism is positively associated with cumulative abnormal stock returns during the COVID-19 crisis.
3. Sample selection and empirical design
3.1 Sample selection
Our sample consists of 2,396 U.S.-listed industrial firms, for which we analyze stock returns over a 45-day window following March 11, 2020, the date when the World Health Organization declared COVID-19 a global pandemic. Each firm is matched to financial statement data from the fiscal year ending in 2019 [2], labeled year t. We exclude firms that, at the end of the year t, have a stock price less than $1, a market value of equity less than $50 million, or total assets of less than $50 million. We also exclude firms with incomplete data necessary for computing the regression variables. The data are retrieved from the Center for Research in Security Prices (CRSP) and the Compustat Fundamentals Annual databases, accessed through the Wharton Research Data Services (WRDS) Website.
3.2 Measures of key variables and estimation methods
3.2.1 Independent variables
To capture the level of conservatism employed by each firm before the crisis, we follow prior research and adopt a composite proxy (CONS) derived as the average of the decile ranks of three conservatism measures commonly used in the literature (e.g. Biddle et al., 2022; Francis et al., 2013; Hui et al., 2009; Jaggi et al., 2021; Mashruwala and Mashruwala, 2018). The first measure (CONS_Res) stems from Penman and Zhang’s (2002) conjecture that conservatism, coupled with expanding investments, leads to the creation of balance sheet reserves. We estimate conservatism reserves annually, following Penman and Zhang (2002), and compute CONS_Res as the average value of reserves over the three years from t-2 to t.
The second measure (CONS_Market) is based on the notion that conservative accounting tends to bias net assets downward, leading to a persistent understatement of book equity relative to market value (Beaver and Ryan, 2005). We construct CONS_Market as the average end-of-year market-to-book ratio over the three years from t-2 to t.
The third measure (CONS_Accr) is based on the persistent recognition of negative non-operating accruals due to conservative accounting practices (Givoly and Hayn, 2000). Non-operating accruals are calculated annually and scaled by lagged total assets, following Givoly and Hayn (2000). We then construct CONS_Accr as the negative of the average non-operating accruals over the three years from t-2 to t. Within each industry, we compute decile ranks of CONS_Accr, CONS_Market, and CONS_Res and then construct the composite measure CONS as the average of the three decile ranks.
3.2.2 Control variables
The control variables include the prior year’s values of net income before extraordinary items scaled by the market value of equity (ROE) and year-over-year sales growth (ΔSALES). We also control for the logarithm of the stock price at the end of the year (PRICE), the logarithm of total assets (SIZE), and total liabilities divided by total assets (LEVERAGE). To account for pre-pandemic performance, we include the cumulative abnormal returns over the (−90, −30) window (PriorRET), as well as abnormal returns cumulated during the fiscal years t and t−1 (FYRET and LagFYRET, respectively) [3].
3.2.3 Dependent variable and regression equation
We calculate the cumulative abnormal returns over the 45 days starting on March 11, 2020 (day-zero, labeled d) [4]. We label this variable CAR(0,45). Daily abnormal returns are the difference between individual stock returns and the value-weighted average market return. We test our hypothesis that conservatism had a beneficial impact on stock market valuations by regressing CAR(0,45) on the conservatism proxy, control variables, and industry fixed effects using the following equation:
In this equation, a positive coefficient estimate for β1 indicates that conservative firms’ abnormal returns outperformed those of less conservative peers following the pandemic’s onset. Tables 1 and 2 report descriptive statistics and pairwise correlations for the regression model variables.
Descriptive statistics
| # obs. | Mean | Std. Dev. | p25 | p50 | p75 | p75 - p25 | |
|---|---|---|---|---|---|---|---|
| CONS_Accr | 1,037 | 0.027 | 0.053 | 0.001 | 0.018 | 0.042 | 0.041 |
| CONS_Market | 2,312 | 4.156 | 4.678 | 1.547 | 2.589 | 4.658 | 3.111 |
| CONS_Res | 1,514 | 0.028 | 0.212 | 0.000 | 0.000 | 0.024 | 0.024 |
| CAR(0,45) | 2,392 | 3.131 | 27.791 | −12.637 | 0.538 | 15.959 | 28.596 |
| RETVOL(0,45) | 2,392 | 3.232 | 2.520 | 1.664 | 2.661 | 4.241 | 2.577 |
| SPREAD(0,45) | 2,392 | 0.234 | 0.497 | 0.020 | 0.085 | 0.228 | 0.208 |
| PriorRET | 2,392 | −0.001 | 0.210 | −0.110 | −0.012 | 0.086 | 0.195 |
| SIZE | 2,392 | 7.578 | 1.841 | 6.200 | 7.506 | 8.802 | 2.602 |
| LEVERAGE | 2,392 | 0.546 | 0.236 | 0.383 | 0.557 | 0.698 | 0.315 |
| ROE | 2,392 | −0.004 | 0.183 | −0.015 | 0.036 | 0.067 | 0.082 |
| PRICE | 2,392 | 3.261 | 1.256 | 2.382 | 3.369 | 4.181 | 1.800 |
| FYRET | 2,392 | −0.013 | 0.479 | −0.302 | −0.056 | 0.182 | 0.485 |
| LagFYRET | 2,392 | −0.026 | 0.379 | −0.280 | −0.079 | 0.151 | 0.431 |
| ΔSALES | 2,392 | 0.094 | 0.180 | 0.004 | 0.055 | 0.139 | 0.135 |
| # obs. | Mean | Std. Dev. | p25 | p50 | p75 | p75 - p25 | |
|---|---|---|---|---|---|---|---|
| CONS_Accr | 1,037 | 0.027 | 0.053 | 0.001 | 0.018 | 0.042 | 0.041 |
| CONS_Market | 2,312 | 4.156 | 4.678 | 1.547 | 2.589 | 4.658 | 3.111 |
| CONS_Res | 1,514 | 0.028 | 0.212 | 0.000 | 0.000 | 0.024 | 0.024 |
| CAR(0,45) | 2,392 | 3.131 | 27.791 | −12.637 | 0.538 | 15.959 | 28.596 |
| RETVOL(0,45) | 2,392 | 3.232 | 2.520 | 1.664 | 2.661 | 4.241 | 2.577 |
| SPREAD(0,45) | 2,392 | 0.234 | 0.497 | 0.020 | 0.085 | 0.228 | 0.208 |
| PriorRET | 2,392 | −0.001 | 0.210 | −0.110 | −0.012 | 0.086 | 0.195 |
| SIZE | 2,392 | 7.578 | 1.841 | 6.200 | 7.506 | 8.802 | 2.602 |
| LEVERAGE | 2,392 | 0.546 | 0.236 | 0.383 | 0.557 | 0.698 | 0.315 |
| ROE | 2,392 | −0.004 | 0.183 | −0.015 | 0.036 | 0.067 | 0.082 |
| PRICE | 2,392 | 3.261 | 1.256 | 2.382 | 3.369 | 4.181 | 1.800 |
| FYRET | 2,392 | −0.013 | 0.479 | −0.302 | −0.056 | 0.182 | 0.485 |
| LagFYRET | 2,392 | −0.026 | 0.379 | −0.280 | −0.079 | 0.151 | 0.431 |
| ΔSALES | 2,392 | 0.094 | 0.180 | 0.004 | 0.055 | 0.139 | 0.135 |
Note(s): This Table reports descriptive statistics for the main variables employed in the regression models. All variables are defined in the appendix
Pearson (lower diagonal) and Spearman (upper diagonal) pairwie correlations
| 1 | 2 | 3 | 4 | 5 | 6 | 7 | 8 | 9 | 10 | 11 | 12 | 13 | 14 | ||
|---|---|---|---|---|---|---|---|---|---|---|---|---|---|---|---|
| 1 | CONS_Accr | 1.000 | |||||||||||||
| 2 | CONS_Market | 0.243* | 1.000 | ||||||||||||
| 3 | CONS_Res | 0.108* | 0.162* | 1.000 | |||||||||||
| 4 | CAR(0,45) | 0.134* | 0.125* | 0.078* | 1.000 | ||||||||||
| 5 | RETVOL(0,45) | 0.091* | −0.160* | −0.042 | 0.233* | 1.000 | |||||||||
| 6 | SPREAD(0,45) | −0.044 | −0.093* | −0.025 | −0.087* | 0.165* | 1.000 | ||||||||
| 7 | PriorRET | 0.142* | 0.152* | 0.035 | 0.119* | −0.104* | −0.035 | 1.000 | |||||||
| 8 | SIZE | −0.211* | −0.049* | −0.024 | −0.138* | −0.253* | −0.368* | −0.140* | 1.000 | ||||||
| 9 | LEVERAGE | 0.069* | 0.237* | −0.051* | −0.138* | 0.126* | −0.086* | −0.044* | 0.324* | 1.000 | |||||
| 10 | ROE | −0.307* | −0.019 | 0.021 | −0.135* | −0.213* | −0.043* | −0.195* | 0.265* | −0.033 | 1.000 | ||||
| 11 | PRICE | −0.044 | 0.294* | 0.073* | −0.087* | −0.485* | −0.246* | 0.006 | 0.466* | 0.017 | 0.324* | 1.000 | |||
| 12 | FYRET | 0.064* | 0.154* | −0.009 | −0.005 | −0.197* | −0.080* | 0.326* | −0.054* | −0.034 | −0.137* | 0.244* | 1.000 | ||
| 13 | LagFYRET | 0.083* | 0.262* | 0.038 | 0.039 | −0.199* | −0.038 | 0.042* | −0.043* | −0.094* | 0.204* | 0.281* | −0.031 | 1.000 | |
| 14 | ΔSALES | 0.104* | 0.154* | 0.096* | 0.046* | 0.006 | −0.001 | −0.026 | −0.045* | −0.011 | 0.180* | 0.109* | −0.049* | 0.162* | 1.000 |
| 1 | 2 | 3 | 4 | 5 | 6 | 7 | 8 | 9 | 10 | 11 | 12 | 13 | 14 | ||
|---|---|---|---|---|---|---|---|---|---|---|---|---|---|---|---|
| 1 | CONS_Accr | 1.000 | |||||||||||||
| 2 | CONS_Market | 0.243* | 1.000 | ||||||||||||
| 3 | CONS_Res | 0.108* | 0.162* | 1.000 | |||||||||||
| 4 | CAR(0,45) | 0.134* | 0.125* | 0.078* | 1.000 | ||||||||||
| 5 | RETVOL(0,45) | 0.091* | −0.160* | −0.042 | 0.233* | 1.000 | |||||||||
| 6 | SPREAD(0,45) | −0.044 | −0.093* | −0.025 | −0.087* | 0.165* | 1.000 | ||||||||
| 7 | PriorRET | 0.142* | 0.152* | 0.035 | 0.119* | −0.104* | −0.035 | 1.000 | |||||||
| 8 | SIZE | −0.211* | −0.049* | −0.024 | −0.138* | −0.253* | −0.368* | −0.140* | 1.000 | ||||||
| 9 | LEVERAGE | 0.069* | 0.237* | −0.051* | −0.138* | 0.126* | −0.086* | −0.044* | 0.324* | 1.000 | |||||
| 10 | ROE | −0.307* | −0.019 | 0.021 | −0.135* | −0.213* | −0.043* | −0.195* | 0.265* | −0.033 | 1.000 | ||||
| 11 | PRICE | −0.044 | 0.294* | 0.073* | −0.087* | −0.485* | −0.246* | 0.006 | 0.466* | 0.017 | 0.324* | 1.000 | |||
| 12 | FYRET | 0.064* | 0.154* | −0.009 | −0.005 | −0.197* | −0.080* | 0.326* | −0.054* | −0.034 | −0.137* | 0.244* | 1.000 | ||
| 13 | LagFYRET | 0.083* | 0.262* | 0.038 | 0.039 | −0.199* | −0.038 | 0.042* | −0.043* | −0.094* | 0.204* | 0.281* | −0.031 | 1.000 | |
| 14 | ΔSALES | 0.104* | 0.154* | 0.096* | 0.046* | 0.006 | −0.001 | −0.026 | −0.045* | −0.011 | 0.180* | 0.109* | −0.049* | 0.162* | 1.000 |
Note(s): This Table reports Pearson pairwise correlations between the main variables employed in the regression models. All variables are defined in the appendix
3.3 Descriptive analysis
Before discussing the results of the empirical models, we provide descriptive analysis by visually examining stock market patterns at the onset of the pandemic. Figure 1 shows the value-weighted average of daily buy-and-hold stock returns from the day d-20 to d+45 for high- and low-conservatism firms, with the median value of CONS serving as the cutoff between the two groups.
The horizontal axis of the line graph is labeled “Day” with values ranging from negative 20 to 40 in increments of 20 units. The vertical axis ranges from negative 40 to 0 percent in increments of 10 percent. Two lines are shown in the graph. A legend at the bottom indicates that the blue line with cross markers represents “High-conservatism firms,” and the red line with circular markers represents “Low-conservatism firms.” Dashed vertical and horizontal lines are drawn from the 0 mark on the horizontal and vertical axes, respectively. The details of both the lines are as follows: The line for “High-conservatism firms” starts at (minus 20, 1), stays steady with small fluctuations till (negative 14, 1.1). It drops steeply to (minus 8, negative 11.0). It rises slightly and again drops to (negative 2, 17.33). It again rises and falls to (1, minus 25.93). It decreases with constant fluctuations and reaches of low of (8, negative 31.07). The line then gradually rises with constant fluctuations to (11, negative 19.60). It drops and continues upward with fluctuations to (23, negative 12.69), (34, negative 7.79), and ends at (45, negative 9.97). The line for “Low-conservatism firms” starts at (minus 20, 1), stays steady with small fluctuations till (negative 13, 0.6). It drops steeply to (minus 8, negative 10.59). It rises slightly and again drops to (negative 2, 18.99). It again rises and falls to (1, minus 28.57). It decreases with constant fluctuations and reaches of low of (8, negative 35.71). The line then gradually rises with constant fluctuations to (11, negative 23.28). It drops and continues upward with fluctuations to (26, negative 17.3), (34, negative 13.39), and falls again to end at (45, negative 19.16). Note: All the numerical values are approximated.Average buy-and-hold returns before and after March 11, 2020. Source: Authors’ own work
The horizontal axis of the line graph is labeled “Day” with values ranging from negative 20 to 40 in increments of 20 units. The vertical axis ranges from negative 40 to 0 percent in increments of 10 percent. Two lines are shown in the graph. A legend at the bottom indicates that the blue line with cross markers represents “High-conservatism firms,” and the red line with circular markers represents “Low-conservatism firms.” Dashed vertical and horizontal lines are drawn from the 0 mark on the horizontal and vertical axes, respectively. The details of both the lines are as follows: The line for “High-conservatism firms” starts at (minus 20, 1), stays steady with small fluctuations till (negative 14, 1.1). It drops steeply to (minus 8, negative 11.0). It rises slightly and again drops to (negative 2, 17.33). It again rises and falls to (1, minus 25.93). It decreases with constant fluctuations and reaches of low of (8, negative 31.07). The line then gradually rises with constant fluctuations to (11, negative 19.60). It drops and continues upward with fluctuations to (23, negative 12.69), (34, negative 7.79), and ends at (45, negative 9.97). The line for “Low-conservatism firms” starts at (minus 20, 1), stays steady with small fluctuations till (negative 13, 0.6). It drops steeply to (minus 8, negative 10.59). It rises slightly and again drops to (negative 2, 18.99). It again rises and falls to (1, minus 28.57). It decreases with constant fluctuations and reaches of low of (8, negative 35.71). The line then gradually rises with constant fluctuations to (11, negative 23.28). It drops and continues upward with fluctuations to (26, negative 17.3), (34, negative 13.39), and falls again to end at (45, negative 19.16). Note: All the numerical values are approximated.Average buy-and-hold returns before and after March 11, 2020. Source: Authors’ own work
In the weeks leading up to the WHO’s declaration of the pandemic, both groups experienced sharp declines as it became clear that the virus could not be contained in foreign countries. However, as the pandemic’s stock market effects unfolded, a divergence between the two sets of firms emerged, reaching nearly seven percentage points by day d+8. As the market began to recover following various fiscal and monetary policy interventions aimed at stabilizing the economy, the performance gap between high- and low-conservatism firms persisted, suggesting that this difference was not simply the result of short-term investor overreaction or underreaction to either group of firms.
We extend our visual analysis to stock return volatility, calculated as the standard deviation of daily abnormal returns over a 5-day rolling window centered on each day, and for the bid-ask spread, calculated as the average daily value of abnormal spread over the same 5-day rolling window. Figures 2 and 3 illustrate patterns similar to Figure 1. Just before day zero, low-conservatism firms begin to exhibit significantly higher volatility and wider spreads, consistent with increased investor uncertainty, which raises trading risk and complicates accurate stock pricing. This difference persists for several weeks following the initial spike, reinforcing that conservatism’s impact is driven by longer-term valuation considerations rather than temporary or isolated mispricing.
The horizontal axis of the line graph is labeled “Day” with values ranging from negative 20 to 40 in increments of 20 units. The vertical axis ranges from 0.01 to 0.06 in increments of 0.01 units. Two lines are shown in the graph. A legend at the bottom indicates that the blue line with cross markers represents “High-conservatism firms,” and the red line with circular markers represents “Low-conservatism firms.” A dashed vertical line is drawn from the 0 mark on the horizontal axis. The details of both lines are as follows: The line for “High-conservatism firms” starts at (negative 20, 0.012), slightly increases to form a broad peak at (negative 8, 0.02). It falls in a concave down manner to (negative 5, 0.017), and rises upward to (negative 1, 0.025). It then steeply increases to (4, 0.049). It shows fluctuation and reaches the topmost point of (7, 0.051). It then starts to decrease, falling steeply to (15, 0.024). It shows continuous fluctuations, and reaches (29, 0.019), forms a broad peak at (34, 0.027), decreases in a concave down manner, decreases to (37, 0.02), falls to (44, 0.016), and rises up to end at (45, 0.023). The line for “Low-conservatism firms” starts at (negative 20, 0.014), slightly increases to form a broad peak at (negative 7, 0.021). It falls in a concave down manner to (negative 5, 0.018), and rises upward to (negative 3, 0.031). It then steeply increases with fluctuations to (4, 0.059). It reaches the topmost point of (5, 0.06). It then starts to decrease, falling steeply to (15, 0.029). It shows continuous fluctuations, and reaches (27, 0.025), forms a broad peak at (35, 0.027), and decreases in a concave down manner to (37, 0.021). Forming a broad peak, it decreases to (44, 0.019), and rises up to end at (45, 0.024). Note: All the numerical values are approximated.Abnormal daily return volatility over a five-day rolling window centered on each day before and after March 11, 2020. Source: Authors’ own work
The horizontal axis of the line graph is labeled “Day” with values ranging from negative 20 to 40 in increments of 20 units. The vertical axis ranges from 0.01 to 0.06 in increments of 0.01 units. Two lines are shown in the graph. A legend at the bottom indicates that the blue line with cross markers represents “High-conservatism firms,” and the red line with circular markers represents “Low-conservatism firms.” A dashed vertical line is drawn from the 0 mark on the horizontal axis. The details of both lines are as follows: The line for “High-conservatism firms” starts at (negative 20, 0.012), slightly increases to form a broad peak at (negative 8, 0.02). It falls in a concave down manner to (negative 5, 0.017), and rises upward to (negative 1, 0.025). It then steeply increases to (4, 0.049). It shows fluctuation and reaches the topmost point of (7, 0.051). It then starts to decrease, falling steeply to (15, 0.024). It shows continuous fluctuations, and reaches (29, 0.019), forms a broad peak at (34, 0.027), decreases in a concave down manner, decreases to (37, 0.02), falls to (44, 0.016), and rises up to end at (45, 0.023). The line for “Low-conservatism firms” starts at (negative 20, 0.014), slightly increases to form a broad peak at (negative 7, 0.021). It falls in a concave down manner to (negative 5, 0.018), and rises upward to (negative 3, 0.031). It then steeply increases with fluctuations to (4, 0.059). It reaches the topmost point of (5, 0.06). It then starts to decrease, falling steeply to (15, 0.029). It shows continuous fluctuations, and reaches (27, 0.025), forms a broad peak at (35, 0.027), and decreases in a concave down manner to (37, 0.021). Forming a broad peak, it decreases to (44, 0.019), and rises up to end at (45, 0.024). Note: All the numerical values are approximated.Abnormal daily return volatility over a five-day rolling window centered on each day before and after March 11, 2020. Source: Authors’ own work
The horizontal axis of the line graph is labeled “Day” with values ranging from negative 20 to 40 in increments of 20 units. The vertical axis ranges from 0.000 to 0.006 in increments of 0.002 units. Two lines are plotted in the graph. A legend at the bottom indicates that the blue line with cross markers represents “High-conservatism firms,” and the red line with circular markers represents “Low-conservatism firms.” A dashed vertical line is drawn from the 0 mark on the horizontal axis. The details of both lines are as follows: The line for “High-conservatism firms” starts at (negative 20, 0.00021), rises gradually and reaches about (negative 7, 0.001), then climbs steeply after day 0, peaking at (6, 0.004). It fluctuates downward to (25, 0.001), continues with smaller fluctuations, and ends at (45, 0.001). The line for “Low-conservatism firms” starts near (negative 20, 0), rises gradually to (negative 10.87, 0.00025), and after day 0 it increases more sharply, passing (3, 0.004), (5, 0.005), and peaks at (7, 0.006). It then declines steadily with fluctuations, reaching about (21, 0.002), and continues downward with small peaks and troughs, ending around (45, 0.001). Note: All the numerical values are approximated.Average daily abnormal bid-ask spread over a five-day rolling window centered on each day before and after March 11, 2020. Source: Authors’ own work
The horizontal axis of the line graph is labeled “Day” with values ranging from negative 20 to 40 in increments of 20 units. The vertical axis ranges from 0.000 to 0.006 in increments of 0.002 units. Two lines are plotted in the graph. A legend at the bottom indicates that the blue line with cross markers represents “High-conservatism firms,” and the red line with circular markers represents “Low-conservatism firms.” A dashed vertical line is drawn from the 0 mark on the horizontal axis. The details of both lines are as follows: The line for “High-conservatism firms” starts at (negative 20, 0.00021), rises gradually and reaches about (negative 7, 0.001), then climbs steeply after day 0, peaking at (6, 0.004). It fluctuates downward to (25, 0.001), continues with smaller fluctuations, and ends at (45, 0.001). The line for “Low-conservatism firms” starts near (negative 20, 0), rises gradually to (negative 10.87, 0.00025), and after day 0 it increases more sharply, passing (3, 0.004), (5, 0.005), and peaks at (7, 0.006). It then declines steadily with fluctuations, reaching about (21, 0.002), and continues downward with small peaks and troughs, ending around (45, 0.001). Note: All the numerical values are approximated.Average daily abnormal bid-ask spread over a five-day rolling window centered on each day before and after March 11, 2020. Source: Authors’ own work
4. Empirical results
The patterns illustrated in Figures 1–3 suggest that conservatism alleviated investors’ concerns about firms’ exposure to risks emerging from the unfolding crisis. To test this hypothesis formally, we estimate the regression specified in Equation (1). The results, presented in Table 3, Column 1, indicate that the conservatism measure (CONS) is significantly positively associated with cumulative abnormal returns over the 45 days following the pandemic’s onset. The coefficient implies that a one-decile increase in conservatism improves stock returns by 1.35%. This magnitude is consistent with the gap between the two patterns shown in Figure 1. This implies that the difference between the top and bottom conservatism deciles accounts for approximately 42% [5] of the CAR(0,45) interquartile range from Table 1. Column 2 results show that this positive association remains robust with the inclusion of control variables and fixed effects, leaving the inference essentially unchanged.
Effect of conservatism on stock market returns
| Dependent Var | CAR(0,45) | CAR(0,45) | ||
|---|---|---|---|---|
| Coeff. | t-stat. | Coeff. | t-stat. | |
| CONS | 1.350*** | (5.30) | 1.382*** | (5.22) |
| PriorRET | 1.941** | (2.38) | ||
| SIZE | −0.687 | (−0.95) | ||
| LEVERAGE | −2.894*** | (−4.00) | ||
| ROE | −2.152** | (−2.03) | ||
| PRICE | −0.712 | (−0.87) | ||
| FYRET | −1.056 | (−1.43) | ||
| LagFYRET | 0.440 | (0.63) | ||
| ΔSALES | 0.665 | (1.00) | ||
| Constant | −3.955** | (−2.55) | −4.209*** | (−2.73) |
| Observations | 2,392 | 2,392 | ||
| Adjusted R-squared | 0.013 | 0.174 | ||
| F-statistic | 28.05 | 9.22 | ||
| Industry fixed effects | NO | YES | ||
| Dependent Var | CAR(0,45) | CAR(0,45) | ||
|---|---|---|---|---|
| Coeff. | t-stat. | Coeff. | t-stat. | |
| CONS | 1.350*** | (5.30) | 1.382*** | (5.22) |
| PriorRET | 1.941** | (2.38) | ||
| SIZE | −0.687 | (−0.95) | ||
| LEVERAGE | −2.894*** | (−4.00) | ||
| ROE | −2.152** | (−2.03) | ||
| PRICE | −0.712 | (−0.87) | ||
| FYRET | −1.056 | (−1.43) | ||
| LagFYRET | 0.440 | (0.63) | ||
| ΔSALES | 0.665 | (1.00) | ||
| Constant | −3.955** | (−2.55) | −4.209*** | (−2.73) |
| Observations | 2,392 | 2,392 | ||
| Adjusted R-squared | 0.013 | 0.174 | ||
| F-statistic | 28.05 | 9.22 | ||
| Industry fixed effects | NO | YES | ||
Note(s): This table reports regression estimates examining the association between conservatism and cumulative abnormal returns over the 45 days following the pandemic’s onset. The results show a robust positive relationship, indicating that higher conservatism significantly improved stock market valuations during this period. All variables are defined in the appendix. Industry fixed-effects are included but not reported for brevity. T-statistics of coefficient estimates are reported in parentheses and are based on robust standard errors and two-tailed tests. To facilitate the interpretation of coefficients, CAR(0,45) is multiplied by 100 and control variables are standardized. Continuous variables are winsorized at the 1 and 99% levels. *, **, and *** define significance at the 10%, 5% and 1% levels respectively
Overall, the findings in Table 3 strongly support the view that conservatism positively impacted stock market valuations at the onset of the pandemic. For robustness, we repeat the analysis using shorter time windows to examine how the effect of conservatism evolved as the crisis unfolded. Additionally, we test each component of the composite conservatism measure, CONS_Accr, CONS_Market, and CONS_Res, instead of the overall measure. Table 4 shows a positive association between CONS and stock returns in all three cumulation windows. Moreover, the coefficients for the three conservatism measures are consistent with the main effect and exhibit comparable magnitudes, indicating that all three components contribute to the overall association.
Alternative conservatism proxies and event windows
| Dependent Var | CAR(0,20) | CAR(0,20) | CAR(0,20) | CAR(0,20) | ||||
|---|---|---|---|---|---|---|---|---|
| Coeff. | t-stat. | Coeff. | t-stat. | Coeff. | t-stat. | Coeff. | t-stat. | |
| CONS_Accr | 0.342* | (1.78) | ||||||
| CONS_Market | 0.409** | (2.47) | ||||||
| CONS_Res | 0.581*** | (4.25) | ||||||
| CONS | 0.605*** | (3.35) | ||||||
| Observations | 1,035 | 2,308 | 1,509 | 2,392 | ||||
| Adjusted R-squared | 0.155 | 0.116 | 0.145 | 0.119 | ||||
| F-statistic | 5.21 | 6.16 | 7.27 | 8.47 | ||||
| FE, constant, controls | YES | YES | YES | YES | ||||
| Dependent Var | CAR(0,20) | CAR(0,20) | CAR(0,20) | CAR(0,20) | ||||
|---|---|---|---|---|---|---|---|---|
| Coeff. | t-stat. | Coeff. | t-stat. | Coeff. | t-stat. | Coeff. | t-stat. | |
| CONS_Accr | 0.342* | (1.78) | ||||||
| CONS_Market | 0.409** | (2.47) | ||||||
| CONS_Res | 0.581*** | (4.25) | ||||||
| CONS | 0.605*** | (3.35) | ||||||
| Observations | 1,035 | 2,308 | 1,509 | 2,392 | ||||
| Adjusted R-squared | 0.155 | 0.116 | 0.145 | 0.119 | ||||
| F-statistic | 5.21 | 6.16 | 7.27 | 8.47 | ||||
| FE, constant, controls | YES | YES | YES | YES | ||||
| Dependent Var | CAR(0,30) | CAR(0,30) | CAR(0,30) | CAR(0,30) | ||||
|---|---|---|---|---|---|---|---|---|
| Coeff. | t-stat. | Coeff. | t-stat. | Coeff. | t-stat. | Coeff. | t-stat. | |
| CONS_Accr | 0.586** | (2.53) | ||||||
| CONS_Market | 0.810*** | (4.24) | ||||||
| CONS_Res | 0.737*** | (4.29) | ||||||
| CONS | 1.155*** | (5.45) | ||||||
| Observations | 1,035 | 2,308 | 1,509 | 2,392 | ||||
| Adjusted R-squared | 0.184 | 0.182 | 0.173 | 0.187 | ||||
| F-statistic | 3.08 | 5.78 | 4.73 | 7.79 | ||||
| FE, constant, controls | YES | YES | YES | YES | ||||
| Dependent Var | CAR(0,30) | CAR(0,30) | CAR(0,30) | CAR(0,30) | ||||
|---|---|---|---|---|---|---|---|---|
| Coeff. | t-stat. | Coeff. | t-stat. | Coeff. | t-stat. | Coeff. | t-stat. | |
| CONS_Accr | 0.586** | (2.53) | ||||||
| CONS_Market | 0.810*** | (4.24) | ||||||
| CONS_Res | 0.737*** | (4.29) | ||||||
| CONS | 1.155*** | (5.45) | ||||||
| Observations | 1,035 | 2,308 | 1,509 | 2,392 | ||||
| Adjusted R-squared | 0.184 | 0.182 | 0.173 | 0.187 | ||||
| F-statistic | 3.08 | 5.78 | 4.73 | 7.79 | ||||
| FE, constant, controls | YES | YES | YES | YES | ||||
| Dependent Var | CAR(0,45) | CAR(0,45) | CAR(0,45) | CAR(0,45) | ||||
|---|---|---|---|---|---|---|---|---|
| Coeff. | t-stat. | Coeff. | t-stat. | Coeff. | t-stat. | Coeff. | t-stat. | |
| CONS_Accr | 0.557* | (1.86) | ||||||
| CONS_Market | 1.150*** | (5.02) | ||||||
| CONS_Res | 0.845*** | (3.95) | ||||||
| CONS | 1.382*** | (5.22) | ||||||
| Observations | 1,035 | 2,308 | 1,509 | 2,392 | ||||
| Adjusted R-squared | 0.132 | 0.174 | 0.168 | 0.174 | ||||
| F-statistic | 3.07 | 7.69 | 4.28 | 9.22 | ||||
| FE, Constant, Controls | YES | YES | YES | YES | ||||
| Dependent Var | CAR(0,45) | CAR(0,45) | CAR(0,45) | CAR(0,45) | ||||
|---|---|---|---|---|---|---|---|---|
| Coeff. | t-stat. | Coeff. | t-stat. | Coeff. | t-stat. | Coeff. | t-stat. | |
| CONS_Accr | 0.557* | (1.86) | ||||||
| CONS_Market | 1.150*** | (5.02) | ||||||
| CONS_Res | 0.845*** | (3.95) | ||||||
| CONS | 1.382*** | (5.22) | ||||||
| Observations | 1,035 | 2,308 | 1,509 | 2,392 | ||||
| Adjusted R-squared | 0.132 | 0.174 | 0.168 | 0.174 | ||||
| F-statistic | 3.07 | 7.69 | 4.28 | 9.22 | ||||
| FE, Constant, Controls | YES | YES | YES | YES | ||||
Note(s): This table reports regression results assessing the robustness of the conservatism–returns association across shorter event windows and for individual conservatism components. The results show consistent positive effects for all windows and components. All variables are defined in the appendix. Industry fixed-effects, the constant term, and control variables from Table 3 are included in all models but not reported for brevity. T-statistics of coefficient estimates are reported in parentheses and are based on robust standard errors and two-tailed tests. To facilitate the interpretation of coefficients, CAR(0,45), CAR(0,30), and CAR(0,20) are multiplied by 100. Continuous variables are winsorized at the 1 and 99% level. *, **, and *** define significance at the 10%, 5% and 1% levels respectively
Next, we extend the analysis to determine whether the effect of conservatism observed in Table 3 persists over the longer term. To do this, we test for return reversals in subsequent trading periods by calculating abnormal returns over the windows (+45, +90) and (+45, +150) and regressing them on CAR(0,45), CONS, and all control variables. If the effect observed in Table 3 had been short-lived and reversed in the post-pandemic period, a significantly negative coefficient for CONS would have been observed. However, the results in Table 5 show that, in both models, the coefficient for CONS is insignificant. This suggests that the positive impact on stock market valuations was concentrated around the initial phase of the pandemic, likely driven by crisis-related investor concerns about firms’ cash flow prospects. Additionally, the results indicate that the effect was not due to temporary mispricing during the crisis, potentially due to lockdowns’ dampening market efficiency (Bhattacharya, 2024), but had lasting implications for firm valuations.
Long-term persistence of conservatism’s effects
| Dependent Var | CAR(45,90) | CAR(45,150) | ||
|---|---|---|---|---|
| Coeff. | t-stat. | Coeff. | t-stat. | |
| CONS | 0.323 | (1.43) | 0.314 | (0.88) |
| CAR(0,45) | 0.928*** | (38.66) | 0.935*** | (27.86) |
| PriorRET | 0.096 | (0.12) | −0.751 | (−0.68) |
| SIZE | 0.763 | (1.20) | −0.187 | (−0.20) |
| LEVERAGE | 1.892*** | (2.97) | 3.555*** | (3.73) |
| ROE | −2.898*** | (−3.11) | −1.823 | (−1.34) |
| PRICE | −5.337*** | (−7.33) | −7.755*** | (−6.72) |
| FYRET | −0.597 | (−0.88) | 1.148 | (1.09) |
| LagFYRET | −0.295 | (−0.52) | 0.346 | (0.37) |
| ΔSALES | 1.136** | (2.14) | 1.757** | (2.09) |
| Constant | 6.781*** | (5.26) | 8.714*** | (4.33) |
| Observations | 2,392 | 2,392 | ||
| Adjusted R-squared | 0.630 | 0.441 | ||
| F-statistic | 186.19 | 99.21 | ||
| Industry fixed effects | YES | YES | ||
| Dependent Var | CAR(45,90) | CAR(45,150) | ||
|---|---|---|---|---|
| Coeff. | t-stat. | Coeff. | t-stat. | |
| CONS | 0.323 | (1.43) | 0.314 | (0.88) |
| CAR(0,45) | 0.928*** | (38.66) | 0.935*** | (27.86) |
| PriorRET | 0.096 | (0.12) | −0.751 | (−0.68) |
| SIZE | 0.763 | (1.20) | −0.187 | (−0.20) |
| LEVERAGE | 1.892*** | (2.97) | 3.555*** | (3.73) |
| ROE | −2.898*** | (−3.11) | −1.823 | (−1.34) |
| PRICE | −5.337*** | (−7.33) | −7.755*** | (−6.72) |
| FYRET | −0.597 | (−0.88) | 1.148 | (1.09) |
| LagFYRET | −0.295 | (−0.52) | 0.346 | (0.37) |
| ΔSALES | 1.136** | (2.14) | 1.757** | (2.09) |
| Constant | 6.781*** | (5.26) | 8.714*** | (4.33) |
| Observations | 2,392 | 2,392 | ||
| Adjusted R-squared | 0.630 | 0.441 | ||
| F-statistic | 186.19 | 99.21 | ||
| Industry fixed effects | YES | YES | ||
Note(s): This table reports regression results testing whether the positive effect of conservatism on stock returns persisted beyond the initial pandemic period. The results show no evidence of return reversals, suggesting that the valuation impact was concentrated at the onset of the crisis and not driven by temporary mispricing. All variables are defined in the appendix. Industry fixed-effects are included but not reported for brevity. T-statitics of coefficient estimates are reported in parentheses and are based on robust standard errors and two-tailed tests. To facilitate the interpretation of coefficients, CAR(0,45), CAR(45,90), and CAR(45,150) are multiplied by 100 and control variables are standardized. Continuous variables are winsorized at the 1 and 99% level. *, **, and *** define significance at the 10%, 5% and 1% levels respectively
Finally, we evaluate whether conservatism reduced investor uncertainty during the pandemic by analyzing daily abnormal return volatility and average abnormal bid-ask spreads. Specifically, we compute the proxy for abnormal return volatility, labeled RETVOL(0,45), as the volatility of daily abnormal returns over the (0,45) window less the volatility over the (−90,−30) window. We compute the proxy for abnormal bid-ask spreads, labeled SPREAD(0,45), as the average daily spread over the (0,45) window less the average spread over the (−90,−30) window.
Table 6 shows that conservatism’s association with both proxies is significantly negative, suggesting that moving from the bottom to the top conservatism decile rank reduces abnormal volatility of daily returns by 0.756% and abnormal spread by 0.261% of the stock price [6]. Consistent with our previous findings, these results indicate that conservatism mitigated the surge in investor uncertainty at the onset of the pandemic, reducing the risk associated with trading the stock.
Effect of conservatism on return volatility and bid-ask spread
| Dependent Var | RETVOL(0,45) | SPREAD(0,45) | ||
|---|---|---|---|---|
| Coeff. | t-stat. | Coeff. | t-stat. | |
| CONS | −0.084*** | (−3.64) | −0.029*** | (−5.54) |
| PriorRET | −0.591*** | (−8.91) | −0.017* | (−1.76) |
| SIZE | −0.286*** | (−5.07) | −0.230*** | (−12.04) |
| LEVERAGE | 0.451*** | (7.28) | 0.025** | (2.03) |
| ROE | −0.062 | (−0.79) | 0.016 | (1.01) |
| PRICE | −0.167** | (−2.51) | 0.011 | (0.63) |
| FYRET | −0.089 | (−1.55) | −0.025** | (−2.17) |
| LagFYRET | −0.050 | (−0.91) | −0.011 | (−0.98) |
| ΔSALES | 0.027 | (0.53) | −0.008 | (−0.64) |
| Constant | 3.720*** | (29.20) | 0.391*** | (12.39) |
| Observations | 2,392 | 2,392 | ||
| Adjusted R-squared | 0.287 | 0.192 | ||
| F-statistic | 26.98 | 30.57 | ||
| Industry fixed effects | YES | YES | ||
| Dependent Var | RETVOL(0,45) | SPREAD(0,45) | ||
|---|---|---|---|---|
| Coeff. | t-stat. | Coeff. | t-stat. | |
| CONS | −0.084*** | (−3.64) | −0.029*** | (−5.54) |
| PriorRET | −0.591*** | (−8.91) | −0.017* | (−1.76) |
| SIZE | −0.286*** | (−5.07) | −0.230*** | (−12.04) |
| LEVERAGE | 0.451*** | (7.28) | 0.025** | (2.03) |
| ROE | −0.062 | (−0.79) | 0.016 | (1.01) |
| PRICE | −0.167** | (−2.51) | 0.011 | (0.63) |
| FYRET | −0.089 | (−1.55) | −0.025** | (−2.17) |
| LagFYRET | −0.050 | (−0.91) | −0.011 | (−0.98) |
| ΔSALES | 0.027 | (0.53) | −0.008 | (−0.64) |
| Constant | 3.720*** | (29.20) | 0.391*** | (12.39) |
| Observations | 2,392 | 2,392 | ||
| Adjusted R-squared | 0.287 | 0.192 | ||
| F-statistic | 26.98 | 30.57 | ||
| Industry fixed effects | YES | YES | ||
Note(s): This table reports regression results examining the association between conservatism and changes in abnormal return volatility and bid-ask spreads during the pandemic onset. The results show that higher conservatism significantly reduced both measures, indicating lower investor uncertainty and trading risk. All variables are defined in the appendix. Industry fixed-effects are included but not reported for brevity. T-statistics of coefficient estimates are reported in parentheses and are based on robust standard errors and two-tailed tests. To facilitate the interpretation of coefficients, RETVOL(0,45) and SPREAD(0,45) are multiplied by 100 and control variables are standardized. Continuous variables are winsorized at the 1 and 99% level. *, **, and *** define significance at the 10%, 5% and 1% levels respectively
5. Additional analyses
In this section, we extend our analysis by examining additional questions that build on the core findings. These supplemental tests provide further insight into the mechanisms underlying our main results and help assess the robustness and broader implications of our conclusions.
5.1 The effect of conservatism as the pandemic worsened from January through April, 2020
In this section, we broaden the scope of the analysis to examine the sequence of escalating events that contributed to the deepening crisis. Figure 4 displays the two-day cumulative abnormal returns of the S&P 500 between January and mid-April 2020. Although the market responded rapidly, the figure shows that the pandemic unfolded as a series of increasingly severe negative shocks beginning in late January and intensifying through early March.
The vertical axis of the vertical bar graph is labeled with values from negative 15 percent to positive 10 percent in increments of 5 percent. The horizontal axis has dates labeled January 1st, February 1st, March 1st, and April 1st. Several blue bars of varying heights extend both upward and downward along the vertical axis. Four text annotations with arrows point to specific bars. The text next to the bar showing 23rd January reads “Lockdown of Wuhan (China), January 23rd.” The text next to the bar showing 23rd February reads “First C O V I D cases and urgent restrictions in Italy, February 21st–24th.” The text next to the bar showing 11th March reads “W H O declares C O V I D-19 a pandemic, March 11th.” The text next to the bar showing 31st March reads “White House announcement of up to 240k expected deaths even with countermeasures, March 31st.” The complete data from the graph is as follows: 1 January: Not given. 3 January: Negative 0.29 percent. 5 January: Not given. 7 January: 0.19 percent. 9 January: 0.41 percent. 11 January: Not given. 13 January: 0.58 percent. 15 January: 1.07 percent. 17 January: 0.14 percent. 19 January: Not given. 21 January: 0.14 percent. 23 January: Negative 2.39 percent. 25 January: Not given. 27 January: 0.91 percent. 29 January: Negative 1.28 percent. 31 January: 2.23 percent. 3 February: 1.46 percent. 7 February: 0.26 percent. 9 February: Not given. 11 February: 0.91 percent. 13 February: 0.88 percent. 15 February: Not given. 17 February: 0.25 percent. 19 February: Negative 1.34 percent. 21 February: Not given. 23 February: Negative 6.3 percent. 25 February: Negative 4.70 percent. 27 February: 3.77 percent. 1 March: Not given. 3 March: 1.41 percent. 5 March: Negative 4.97 percent. 7 March: Not given. 9 March: Negative 2.5 percent. 11 March: Negative 14.28 percent. 13 March: Negative 2.5 percent. 15 March: Not given. 17 March: 0.96 percent. 19 March: Negative 3.76 percent. 21 March: Not given. 23 March: 6.42 percent. 25 March: 7.35 percent. 27 March: 0.88 percent. 29 March: Not given. 31 March: Negative 5.85 percent. 2 April: 0.74 percent. 4 April: Not given. 6 April: 6.80 percent. 8 April: 4.82 percent. 10 April: Not given. 12 April: 2.18 percent. 14 April: negative 1.45 16 April: 0.85 percent. 18 April: Not given. 20 April: negative 0.73 percent. 20 April: 1.35 percent. Note: All numerical values are approximated.Two-day cumulated returns of the S&P 500 index in the first months of 2020. Source: Authors’ own work
The vertical axis of the vertical bar graph is labeled with values from negative 15 percent to positive 10 percent in increments of 5 percent. The horizontal axis has dates labeled January 1st, February 1st, March 1st, and April 1st. Several blue bars of varying heights extend both upward and downward along the vertical axis. Four text annotations with arrows point to specific bars. The text next to the bar showing 23rd January reads “Lockdown of Wuhan (China), January 23rd.” The text next to the bar showing 23rd February reads “First C O V I D cases and urgent restrictions in Italy, February 21st–24th.” The text next to the bar showing 11th March reads “W H O declares C O V I D-19 a pandemic, March 11th.” The text next to the bar showing 31st March reads “White House announcement of up to 240k expected deaths even with countermeasures, March 31st.” The complete data from the graph is as follows: 1 January: Not given. 3 January: Negative 0.29 percent. 5 January: Not given. 7 January: 0.19 percent. 9 January: 0.41 percent. 11 January: Not given. 13 January: 0.58 percent. 15 January: 1.07 percent. 17 January: 0.14 percent. 19 January: Not given. 21 January: 0.14 percent. 23 January: Negative 2.39 percent. 25 January: Not given. 27 January: 0.91 percent. 29 January: Negative 1.28 percent. 31 January: 2.23 percent. 3 February: 1.46 percent. 7 February: 0.26 percent. 9 February: Not given. 11 February: 0.91 percent. 13 February: 0.88 percent. 15 February: Not given. 17 February: 0.25 percent. 19 February: Negative 1.34 percent. 21 February: Not given. 23 February: Negative 6.3 percent. 25 February: Negative 4.70 percent. 27 February: 3.77 percent. 1 March: Not given. 3 March: 1.41 percent. 5 March: Negative 4.97 percent. 7 March: Not given. 9 March: Negative 2.5 percent. 11 March: Negative 14.28 percent. 13 March: Negative 2.5 percent. 15 March: Not given. 17 March: 0.96 percent. 19 March: Negative 3.76 percent. 21 March: Not given. 23 March: 6.42 percent. 25 March: 7.35 percent. 27 March: 0.88 percent. 29 March: Not given. 31 March: Negative 5.85 percent. 2 April: 0.74 percent. 4 April: Not given. 6 April: 6.80 percent. 8 April: 4.82 percent. 10 April: Not given. 12 April: 2.18 percent. 14 April: negative 1.45 16 April: 0.85 percent. 18 April: Not given. 20 April: negative 0.73 percent. 20 April: 1.35 percent. Note: All numerical values are approximated.Two-day cumulated returns of the S&P 500 index in the first months of 2020. Source: Authors’ own work
The first notable shock occurred on January 23, when Wuhan, a city of 11 million people in China, was placed under lockdown, prompting the S&P 500 to drop nearly 2.5% over two days. A second, more substantial shock took place over the weekend of February 22–24, as a sudden outbreak in northern Italy prompted urgent restrictions and dashed hopes that the virus could be locally contained. In the days that followed, the market shed over 10%. The largest shock of the period occurred when the World Health Organization officially declared COVID-19 a global pandemic on March 11, leading to a two-day market loss of nearly 15%. A final significant shock came about three weeks later, when a White House press briefing projected between 100,000 and 240,000 potential deaths in the U.S., even under strict public health measures, triggering another 5% market decline.
While the main analysis focused on the largest shock, i.e. March 11, this section replicates the analysis alternatively using the three other significant shocks identified earlier as the starting point of the return cumulation window. The results, presented in Table 7, Panel A, consistently indicate that firms with higher levels of conservatism experienced better stock returns as the crisis unfolded. These findings reinforce the study’s main inference: conservatism’s protective effect is evident throughout the crisis period and is not sensitive to choosing any specific event as the starting point of the cumulation window.
The effect of conservatism as the pandemic worsened from January through April, 2020, and the difference between firms with high and low conservatism reserves at the beginning of the year
| Dependent Var | CAR | CAR | CAR | |||
|---|---|---|---|---|---|---|
| First day of the cumulation window | January 23rd | Februry 24th | March 31st | |||
| Coeff. | t-stat. | Coeff. | t-stat. | Coeff. | t-stat. | |
| Panel A: Full sample | ||||||
| CONS | 2.083*** | (6.47) | 1.701*** | (5.71) | 0.872*** | (3.62) |
| Observations | 2,392 | 2,392 | 2,392 | |||
| Adjusted R-squared | 0.221 | 0.199 | 0.180 | |||
| F-statistic | 16.77 | 14.37 | 7.42 | |||
| Constant and Control variables | YES | YES | YES | |||
| Industry fixed effects | YES | YES | YES | |||
| Panel B: Subsample of firms with smaller conservatism reserves at the onset of the year | ||||||
| CONS | 1.443* | (1.69) | 0.592 | (0.74) | −0.246 | (−0.41) |
| Observations | 1,249 | 1,249 | 1,249 | |||
| Adjusted R-squared | 0.165 | 0.156 | 0.227 | |||
| F-statistic | 4.51 | 6.24 | 3.64 | |||
| Constant and Control variables | YES | YES | YES | |||
| Industry fixed effects | YES | YES | YES | |||
| Panel C: Subsample of firms with larger conservatism reserves at the onset of the year | ||||||
| CONS | 2.286*** | (3.29) | 2.030*** | (3.28) | 1.613*** | (3.24) |
| Observations | 1,136 | 1,136 | 1,136 | |||
| Adjusted R-squared | 0.274 | 0.259 | 0.138 | |||
| F-statistic | 8.94 | 6.54 | 5.55 | |||
| Constant and control variables | YES | YES | YES | |||
| Industry fixed effects | YES | YES | YES | |||
| Difference in the CONS coefficient between Panel B and Panel C | Diff. | p-value | Diff. | p-value | Diff. | p-value |
| 0.843 | [0.445] | 1.438 | [0.157] | 1.859** | [0.017] | |
| Dependent Var | CAR | CAR | CAR | |||
|---|---|---|---|---|---|---|
| First day of the cumulation window | January 23rd | Februry 24th | March 31st | |||
| Coeff. | t-stat. | Coeff. | t-stat. | Coeff. | t-stat. | |
| Panel A: Full sample | ||||||
| CONS | 2.083*** | (6.47) | 1.701*** | (5.71) | 0.872*** | (3.62) |
| Observations | 2,392 | 2,392 | 2,392 | |||
| Adjusted R-squared | 0.221 | 0.199 | 0.180 | |||
| F-statistic | 16.77 | 14.37 | 7.42 | |||
| Constant and Control variables | YES | YES | YES | |||
| Industry fixed effects | YES | YES | YES | |||
| Panel B: Subsample of firms with smaller conservatism reserves at the onset of the year | ||||||
| CONS | 1.443* | (1.69) | 0.592 | (0.74) | −0.246 | (−0.41) |
| Observations | 1,249 | 1,249 | 1,249 | |||
| Adjusted R-squared | 0.165 | 0.156 | 0.227 | |||
| F-statistic | 4.51 | 6.24 | 3.64 | |||
| Constant and Control variables | YES | YES | YES | |||
| Industry fixed effects | YES | YES | YES | |||
| Panel C: Subsample of firms with larger conservatism reserves at the onset of the year | ||||||
| CONS | 2.286*** | (3.29) | 2.030*** | (3.28) | 1.613*** | (3.24) |
| Observations | 1,136 | 1,136 | 1,136 | |||
| Adjusted R-squared | 0.274 | 0.259 | 0.138 | |||
| F-statistic | 8.94 | 6.54 | 5.55 | |||
| Constant and control variables | YES | YES | YES | |||
| Industry fixed effects | YES | YES | YES | |||
| Difference in the CONS coefficient between Panel B and Panel C | Diff. | p-value | Diff. | p-value | Diff. | p-value |
| 0.843 | [0.445] | 1.438 | [0.157] | 1.859** | [0.017] | |
Note(s): This table reports regression results examining whether the positive effect of conservatism on stock returns persisted across multiple major pandemic-related shocks from January to April 2020. The results show a consistent protective effect across events, with evidence that firms with higher initial reserves maintained this advantage longer than firms with lower reserves. All variables are defined in the appendix. Industry fixed-effects, the constant term, and control variables from Table 3 are included but not reported for brevity. T-statistics of coefficient estimates are reported in parentheses and are based on robust standard errors and two-tailed tests. P-values of coefficient differences across models are reported in brakets and are based on robust standard errors and two-tailed tests. To facilitate the interpretation of coefficients, CAR is multiplied by 100 and control variables are standardized. In each of the three models, the first day of the return cumulation window varies (i.e. January 23, Febryary 24, and March 31) while the last day is the same (i.e. the 45th trading day after March 11). Continuous variables are winsorized at the 1 and 99% level. *, **, and *** define significance at the 10%, 5% and 1% levels respectively
The results in Table 7, Panel A, also reveal a declining trend in the coefficient of CONS as the crisis unfolds, potentially suggesting a gradual weakening of conservatism’s beneficial impact on the stock market. If unconditional conservatism builds balance sheet reserves that help support valuations during periods of stress, those reserves may be drawn down over the course of repeated negative shocks. This effect is likely to be more pronounced for firms that entered the crisis with lower levels of conservatism, that is, with fewer pre-existing reserves. For these firms, conservatism may have had a stronger positive effect on stock returns during the early stages of the crisis, but its influence would diminish as the pandemic worsened and shocks accumulated. By contrast, firms with higher initial reserves would be better positioned to sustain valuations throughout the crisis period, allowing conservatism’s beneficial effects to emerge over prolonged periods.
To test this idea [7], we re-estimate the model after splitting the sample based on firms’ conservatism levels at the start of 2020, using the median CONS value as the cutoff. Panels B and C of Table 7 show that when the full crisis period is considered (including the initial shocks in late January) the CONS coefficient is not statistically different across the high- and low-reserve groups. However, a divergence appears as the analysis window narrows to focus on the later stages of the crisis. Specifically, for firms with low reserves, the CONS coefficient steadily declines in both magnitude and significance, eventually becoming negligible when only March is considered. This pattern supports the idea that the effect of conservatism is quickly depleted when reserves are limited. In contrast, firms with high initial reserves continue to exhibit a strong positive market response, suggesting that greater conservatism enables sustained investor confidence over a longer series of shocks.
5.2 Financial distress, operating cash flow volatility, and reporting transparency
The results from the main analysis suggest that conservatism alleviated investors’ concerns about firm prospects as the pandemic broke out. This section explores whether conservatism’s impact varies across firm characteristics that may amplify or mitigate these concerns.
First, we focus on a setting where the heightened downside risk associated with potential insolvency and litigation triggered by the pandemic is particularly acute. Specifically, we identify firms that recently experienced severe financial stress by calculating the quintile rank of each firm’s Altman’s Z-Score over the prior three years within each industry. We then create the binary variable Fin.Distress, identifying observations in the most distressed quintile, and interact it with CONS in equation (1). The results in Table 8, Column 1, show that, in this specification, the coefficient for CONS*Fin.Distress is significantly positive, indicating that conservatism’s effect is strongest among the most distressed firms where investor concerns about insolvency and downside risk are most justified.
Additional analysis: the moderating effect of financial distress, operating cash flow stability, and reporting transparency
| Dependent Var | CAR(0,45) | CAR(0,45) | CAR(0,45) | |||
|---|---|---|---|---|---|---|
| Coeff. | t-stat. | Coeff. | t-stat. | Coeff. | t-stat. | |
| CONS | 0.905*** | (2.84) | 1.824*** | (5.95) | 1.949*** | (4.24) |
| CONS*Fin.Distress | 1.700** | (2.26) | ||||
| Fin.Distress | −9.004** | (−2.25) | ||||
| CONS*CF.Stability | −1.301** | (−2,13) | ||||
| Fund.Stability | 7.345** | (2.06) | ||||
| CONS*Rep.Transparency | −1.091* | (−1.77) | ||||
| Rep.Transparency | 4.261 | (1.15) | ||||
| Observations | 1970 | 2,208 | 1,438 | |||
| Adjusted R-squared | 0.189 | 0.185 | 0.172 | |||
| F-statistic | 6.18 | 8.35 | 4.84 | |||
| Constant and control variables | YES | YES | YES | |||
| Industry fixed effects | YES | YES | YES | |||
| Dependent Var | CAR(0,45) | CAR(0,45) | CAR(0,45) | |||
|---|---|---|---|---|---|---|
| Coeff. | t-stat. | Coeff. | t-stat. | Coeff. | t-stat. | |
| CONS | 0.905*** | (2.84) | 1.824*** | (5.95) | 1.949*** | (4.24) |
| CONS*Fin.Distress | 1.700** | (2.26) | ||||
| Fin.Distress | −9.004** | (−2.25) | ||||
| CONS*CF.Stability | −1.301** | (−2,13) | ||||
| Fund.Stability | 7.345** | (2.06) | ||||
| CONS*Rep.Transparency | −1.091* | (−1.77) | ||||
| Rep.Transparency | 4.261 | (1.15) | ||||
| Observations | 1970 | 2,208 | 1,438 | |||
| Adjusted R-squared | 0.189 | 0.185 | 0.172 | |||
| F-statistic | 6.18 | 8.35 | 4.84 | |||
| Constant and control variables | YES | YES | YES | |||
| Industry fixed effects | YES | YES | YES | |||
Note(s): This table reports regression results testing whether the effect of conservatism on stock returns during the pandemic varies by financial distress, operating cash flow stability, and reporting transparency. The results show that conservatism’s protective impact is strongest for financially distressed firms, weaker for firms with stable cash flows, and reduced for firms with higher reporting transparency. All variables are defined in the appendix. Industry fixed-effects, the constant term, and control variables from Table 3 are included but not reported for brevity. T-statitics of coefficient estimates are reported in parentheses and are based on robust standard errors and two-tailed tests. To facilitate the interpretation of coefficients, CAR(0,45) is multiplied by 100 and control variables are standardized. Continuous variables are winsorized at the 1 and 99% level. *, **, and *** define significance at the 10%, 5% and 1% levels respectively
Second, we examine the moderating role of fundamental volatility. Firms operating in more stable and predictable business environments are generally less exposed to severe downturns when macroeconomic conditions deteriorate. As a result, the potential for unconditional conservatism to cushion negative shocks may be more limited in these cases. As fundamental volatility grows, firms face greater uncertainty and are more vulnerable to economic stress, making them more likely to benefit from the protective effects of conservatism. For these firms, the presence of reserves created by conservative accounting may help absorb the impact of worsening conditions, thereby supporting stock performance during the crisis.
To identify firms operating in stable and predictable businesses, we look at the volatility of operating cash flows over the prior 16 quarters and create a binary variable (CF.Stability) equal to one for observations in the bottom quintile within the industry. We expect the coefficient of CONS to be weaker for these firms compared to others that operate in less stable and predictable environments. Table 8, Column 2, confirms this expectation. The coefficient on CF.Stability is positive, indicating that investors value stability during times of turmoil. However, its interaction with CONS is negative, suggesting that the role of conservatism in cushioning stock price declines was diminished for firms with more predictable cash flows.
Finally, we examine the potential moderating effect of financial reporting transparency. For more transparent firms, the impact of COVID-19 may be more clearly communicated and understood by investors, reducing uncertainty and providing greater clarity around the appropriate pricing of bad news. This improved information environment can diminish the scope for unconditional conservatism to add value, as investors may already be pricing in risks more precisely. In contrast, as financial reporting becomes more opaque, firms become harder to value, especially in the face of macroeconomic uncertainty. In such cases, investors may apply larger or less targeted discounts, making these firms more likely to benefit from the stabilizing role provided by unconditional conservatism during negative shocks.
We assess financial reporting transparency through the standard deviation of the residuals of the Dechow and Dichev (2002) accrual prediction model over the prior eight years. We create a binary variable (Rep.Transparency) equal to one if such standard deviation is below the industry median and interact it with CONS in the regression model. Table 8, Column 3, shows that the interaction term is negative and significant at the 10% level, consistent with reporting transparency reducing the valuation benefits of conservatism during the COVID-19 crisis.
5.3 The substitutive effect of board independence
In this section, we investigate whether the effect of conservatism substitutes for another corporate governance mechanism known to have beneficial market effects during periods of instability: board independence. Research shows that the board’s role as a key corporate governance tool to protect shareholder interests becomes especially important during turbulent market conditions (Francis et al., 2012). Board independence can lead to more cautious risk-taking following financial downturns (Vallascas et al., 2017), increase the firm’s efforts to regain legitimacy after a crisis event such as a product recall (Hossain Sikdar and Raina, 2024), and influence stakeholders’ perception of organizational crisis management effectiveness (Abebe and Simpson, 2015). Consistent with this picture, Abu Bakar et al. (2020) found that board independence makes creditors more willing to trust the board’s monitoring role, and Muttakin and Khan (2023) showed it has a beneficial effect on integrated reporting quality. Also, in the context of corporate social responsibility (CSR), Jizi et al. (2014) and Saleh and Jurdi (2023) reported that board independence was associated with better CSR performance and disclosure practices, and Gerged et al. (2024) showed that it constrains managerial opportunism in CSR engagement. Based on these insights from prior research, we argue that board independence, by providing a monitoring mechanism that reassures investors about the firm’s governance during the crisis, can substitute for conservatism in alleviating stock market uncertainty and mitigating adverse price reactions to the pandemic.
We use three attributes of board independence: the number of independent directors on the board (#IND), the percentage of independent directors (%IND), and a dummy variable identifying whether the Chairperson is independent (IND_CHAIR). Each of these is interacted with CONS in Equation (1). Table 9 results show that the interaction terms are significantly negative, indicating that the effect of conservatism is weaker when strong board independence increases the quality of the firm’s governance. In contrast, the effect is stronger when there is concern over inadequate monitoring of executive decisions. Overall, these findings support the notion that conservatism acts as a corporate governance mechanism by alleviating investors’ worries during times of crisis.
Additional analysis: the moderating effect of board independence
| Dependent Var | CAR(0,45) | CAR(0,45) | CAR(0,45) | |||
|---|---|---|---|---|---|---|
| Coeff. | t-stat. | Coeff. | t-stat. | Coeff. | t-stat. | |
| CONS | 2.667*** | (6.02) | 2.740*** | (6.40) | 2.411*** | (5.65) |
| CONS*#IND | −0.679*** | (−2.94) | ||||
| CONS*%IND | −0.765*** | (−3.34) | ||||
| CONS*IND_CHAIR | −1.079** | (−1.99) | ||||
| #IND | 2.567* | (1.73) | ||||
| %IND | 2.679* | (1.91) | ||||
| DUAL | 6.376* | (1.93) | ||||
| Observations | 1902 | 1902 | 1902 | |||
| Adjusted R-squared | 0.202 | 0.204 | 0.199 | |||
| F-statistic | 8.15 | 9.05 | 7.68 | |||
| Constant and control variables | YES | YES | YES | |||
| Industry fixed effects | YES | YES | YES | |||
| Dependent Var | CAR(0,45) | CAR(0,45) | CAR(0,45) | |||
|---|---|---|---|---|---|---|
| Coeff. | t-stat. | Coeff. | t-stat. | Coeff. | t-stat. | |
| CONS | 2.667*** | (6.02) | 2.740*** | (6.40) | 2.411*** | (5.65) |
| CONS*#IND | −0.679*** | (−2.94) | ||||
| CONS*%IND | −0.765*** | (−3.34) | ||||
| CONS*IND_CHAIR | −1.079** | (−1.99) | ||||
| #IND | 2.567* | (1.73) | ||||
| %IND | 2.679* | (1.91) | ||||
| DUAL | 6.376* | (1.93) | ||||
| Observations | 1902 | 1902 | 1902 | |||
| Adjusted R-squared | 0.202 | 0.204 | 0.199 | |||
| F-statistic | 8.15 | 9.05 | 7.68 | |||
| Constant and control variables | YES | YES | YES | |||
| Industry fixed effects | YES | YES | YES | |||
Note(s): This table reports regression results testing whether board independence substitutes for the effect of conservatism on stock returns during the pandemic. The results show that conservatism’s impact is weaker when board independence is stronger, consistent with both mechanisms serving similar governance functions in reassuring investors. All variables are defined in the appendix. Industry fixed-effects, the constant term, and control variables from Table 3 are included but not reported for brevity. T-statistics of coefficient estimates are reported in parentheses and are based on robust standard errors and two-tailed tests. To facilitate the interpretation of coefficients, CAR(0,45) is multiplied by 100 and control variables are standardized. Continuous variables are winsorized at the 1 and 99% level. *, **, and *** define significance at the 10%, 5% and 1% levels respectively
5.4 Firm exposure to international markets
We examine whether the effect of conservatism during the COVID-19 crisis varied based on firms’ geographic exposure. As the crisis unfolded, investors gradually came to grasp the global scale of the pandemic and its far-reaching impact on international supply chains and global demand. In this context, purely domestic U.S. firms (i.e. those deriving the majority of their revenues from within the United States) faced comparatively fewer risks stemming from the international spread of the virus. As a result, these firms may have experienced a more muted downside and, correspondingly, a weaker need for the protective effect offered by unconditional conservatism. In contrast, firms with greater exposure to international markets were more vulnerable to the economic disruptions triggered by the pandemic. For these firms, conservatism may have played a more pronounced role in supporting valuations by helping absorb the heightened uncertainty and expected losses linked to global economic conditions.
To explore this possibility, we calculate the percentage of domestic sales over total sales for each observation. We then construct a binary variable (Domestic_Firm) equal to one if the firm’s domestic sales percentage falls within the top quintile of its industry and zero otherwise. As shown in Table 10, the coefficient on Domestic_Firm is significantly positive, while the coefficient on the interaction term CONS*Domestic_Firm is negative. These results are consistent with the idea that firms more exposed to international markets experienced greater valuation pressure during the crisis, thereby creating more scope for conservatism to help sustain their stock prices. In contrast, domestic firms were more insulated from global disruptions and, therefore, had less need for conservatism to play a stabilizing role in sustaining valuations.
Additional analysis: domestic firms vs. other firms
| Dependent Var | CAR(0,45) | |
|---|---|---|
| Coeff. | t-stat. | |
| CONS | 2.190*** | (6.34) |
| CONS*Domestic_Firm | −1.133* | (−1.82) |
| Domestic_Firm | 8.155** | (2.19) |
| Observations | 1943 | |
| Adjusted R-squared | 0.165 | |
| F-statistic | 8.14 | |
| Constant and control variables | YES | |
| Industry fixed effects | YES | |
| Dependent Var | CAR(0,45) | |
|---|---|---|
| Coeff. | t-stat. | |
| CONS | 2.190*** | (6.34) |
| CONS*Domestic_Firm | −1.133* | (−1.82) |
| Domestic_Firm | 8.155** | (2.19) |
| Observations | 1943 | |
| Adjusted R-squared | 0.165 | |
| F-statistic | 8.14 | |
| Constant and control variables | YES | |
| Industry fixed effects | YES | |
Note(s): This table reports regression results testing whether the effect of conservatism on stock returns during the pandemic varied with firms’ exposure to international markets. The results show that conservatism’s impact was stronger for globally exposed firms, while domestic firms experienced less need for its stabilizing role. All variables are defined in the appendix. Industry fixed-effects, the constant term, and control variables from Table 3 are included but not reported for brevity are included but not reported for brevity. T-statistics of coefficient estimates are reported in parentheses and are based on robust standard errors and two-tailed tests. To facilitate the interpretation of coefficients, CAR(0,45) is multiplied by 100 and control variables are standardized. Continuous variables are winsorized at the 1 and 99% levels. *, **, and *** define significance at the 10%, 5% and 1% levels respectively
5.5 Industry exposure to COVID-19
We further examine whether the effect of conservatism varied across industries, depending on their relative exposure to the economic disruptions caused by the pandemic. The impact of COVID-19 was not uniform across sectors. Some industries faced more severe operational and financial challenges due to their heightened sensitivity to macroeconomic shocks, while others were comparatively less affected. For firms in less affected industries, the role of conservatism in influencing investor responses during the crisis may have been more limited.
To investigate this idea, we draw on the measure developed by Hassan et al. (2020), who use textual analysis of quarterly earnings calls to quantify exposure to COVID-19 and other infectious diseases. We compute the average value of Hassan et al.’s exposure proxy for each industry across all four quarters of 2020. We then construct a binary variable (Less_Exposed_Industry), equal to one for industries in the bottom exposure quintile and zero otherwise. As shown in Table 11, the interaction term CONS*Less_Exposed_Industry is significantly negative, indicating that conservatism played a smaller role in supporting stock returns within industries less affected by the virus’s global spread.
Additional analysis: Industries with lower exposure to Covid - 19 vs. other industries
| Dependent Var | CAR(0,45) | |
|---|---|---|
| Coeff. | t-stat. | |
| CONS | 1.654*** | (5.47) |
| CONS*Less_Exposed_Industry | −1.682** | (−2.38) |
| Observations | 2085 | |
| Adjusted R-squared | 0.178 | |
| F-statistic | 8.59 | |
| Constant and control variables | YES | |
| Industry fixed effects | YES | |
| Dependent Var | CAR(0,45) | |
|---|---|---|
| Coeff. | t-stat. | |
| CONS | 1.654*** | (5.47) |
| CONS*Less_Exposed_Industry | −1.682** | (−2.38) |
| Observations | 2085 | |
| Adjusted R-squared | 0.178 | |
| F-statistic | 8.59 | |
| Constant and control variables | YES | |
| Industry fixed effects | YES | |
Note(s): This table reports regression results testing whether the effect of conservatism on stock returns during the pandemic varied by industry exposure to COVID-19. The results show that conservatism’s influence was weaker in less exposed industries, consistent with its reduced importance when operational and financial disruptions were limited. All variables are defined in the appendix. Industry fixed-effects, the constant term, and control variables from Table 3 are included but not reported for brevity The coefficient of the Less_Exposed_Industry identifier is absorbed by industry fixed-effects. T-statistics of coefficient estimates are reported in parentheses and are based on robust standard errors and two-tailed tests. To facilitate the interpretation of coefficients, CAR(0,45) is multiplied by 100 and control variables are standardized. Continuous variables are winsorized at the 1 and 99% levels. *, **, and *** define significance at the 10%, 5% and 1% levels respectively
5.6 Special items
In this section, we analyze the financial statements from year t+1 (the fiscal year underway when the pandemic struck) to assess whether investors were justified in their concerns about the exposure of less conservative firms to losses caused by COVID-19. If the crisis heightened uncertainty around asset valuations, triggering fears of imminent impairments and write-downs, these appeared in subsequent income statements. However, the conservative valuation of net assets made negative special items less likely, thereby limiting their impact on earnings performance.
To test this expectation, we analyze the association between CONS and the special items reported in the t+1 income statement (SPIt+1), scaled by the beginning market value of equity. The results in Table 12 show that regressing SPIt+1 on CONS yields a positive coefficient [8], indicating that special items were less negative among conservative firms than others.
Additional analysis: the effect of conservatism on 2020 negative special items
| Dependent Var | SPI(t+1) | |
|---|---|---|
| Coeff. | t-stat. | |
| CONS | 0.030*** | (2.70) |
| PriorRET | 0.011 | (0.34) |
| SIZE | −0.157*** | (−5.14) |
| LEVERAGE | −0.046* | (−1.81) |
| ROE | 0.051 | (1.02) |
| PRICE | 0.216*** | (5.79) |
| FYRET | 0.118*** | (4.29) |
| LagFYRET | 0.073*** | (3.26) |
| ΔSALES | −0.035 | (−1.41) |
| Constant | −0.426*** | (−6.54) |
| Observations | 2,266 | |
| Adjusted R-squared | 0.104 | |
| F-statistic | 12.15 | |
| Industry fixed effects | YES | |
| Dependent Var | SPI(t+1) | |
|---|---|---|
| Coeff. | t-stat. | |
| CONS | 0.030*** | (2.70) |
| PriorRET | 0.011 | (0.34) |
| SIZE | −0.157*** | (−5.14) |
| LEVERAGE | −0.046* | (−1.81) |
| ROE | 0.051 | (1.02) |
| PRICE | 0.216*** | (5.79) |
| FYRET | 0.118*** | (4.29) |
| LagFYRET | 0.073*** | (3.26) |
| ΔSALES | −0.035 | (−1.41) |
| Constant | −0.426*** | (−6.54) |
| Observations | 2,266 | |
| Adjusted R-squared | 0.104 | |
| F-statistic | 12.15 | |
| Industry fixed effects | YES | |
Note(s): This table reports regression results testing the association between conservatism and special items reported in the year following the pandemic’s onset. The results show that conservative firms recorded less negative special items, consistent with their lower exposure to asset impairments and write-downs. All variables are defined in the appendix. Industry fixed-effects are included but not reported for brevity. T-statistics of coefficient estimates are reported in parentheses and are based on robust standard errors and two-tailed tests. To facilitate the interpretation of coefficients, SPI(t+1) and control variables are standardized. Continuous variables are winsorized at the 1 and 99% level. *, **, and *** define significance at the 10%, 5% and 1% levels respectively
5.7 The preemptive effect of unconditional conservatism on conditional conservatism in 2020 income statements
The results in Table 12 indicate that unconditionally conservative reserves helped shield firms from recording impairments and write-downs during the crisis, supporting the idea of a substitution effect between conditional and unconditional conservatism. Firms with higher unconditional conservatism faced fewer or less severe negative earnings shocks, reducing the need for conditional recognition of bad news in the income statement. To further investigate this substitution effect, we conduct a supplemental analysis examining whether the level of unconditional conservatism at the start of 2020 (proxied by CONS) is negatively associated with the degree of conditional conservatism in 2020 income statements (measured by the asymmetric timeliness coefficient of RET*D.RET in the Basu (1997) model). As shown in Table 13, the coefficient on RET*D.RET*CONS is significantly negative, consistent with the presence of a substitution relationship between the two forms of conservatism.
Additional analysis: the relationship between conditional and unconditional conservatism in 2020
| Dependent Var | EARN | EARN | ||
|---|---|---|---|---|
| Coeff. | t-stat. | Coeff. | t-stat. | |
| RET | −0.037 | (−0.91) | 0.025 | (0.69) |
| D.RET | 0.158*** | (3.24) | 0.098** | (2.43) |
| RET*D.RET | 0.676*** | (6.20) | 0.408*** | (4.47) |
| CONS | 0.010* | (1.83) | 0.000 | (0.05) |
| RET*CONS | 0.000 | (0.03) | −0.004 | (−0.62) |
| D.RET*CONS | −0.021*** | (−2.80) | −0.014** | (−2.11) |
| RET*D.RET*CONS | −0.067*** | (−3.80) | −0.050*** | (−3.14) |
| Control Variables | YES | YES | ||
| RET * Control Variables | NO | YES | ||
| D.RET * Control Variables | NO | YES | ||
| RET * D.RET * Control Variables | NO | YES | ||
| Constant and Ind. Fixed Effects | YES | YES | ||
| Observations | 2,284 | 2,284 | ||
| Adjusted R-squared | 0.225 | 0.351 | ||
| F-statistic | 19.24 | 10.82 | ||
| Dependent Var | EARN | EARN | ||
|---|---|---|---|---|
| Coeff. | t-stat. | Coeff. | t-stat. | |
| RET | −0.037 | (−0.91) | 0.025 | (0.69) |
| D.RET | 0.158*** | (3.24) | 0.098** | (2.43) |
| RET*D.RET | 0.676*** | (6.20) | 0.408*** | (4.47) |
| CONS | 0.010* | (1.83) | 0.000 | (0.05) |
| RET*CONS | 0.000 | (0.03) | −0.004 | (−0.62) |
| D.RET*CONS | −0.021*** | (−2.80) | −0.014** | (−2.11) |
| RET*D.RET*CONS | −0.067*** | (−3.80) | −0.050*** | (−3.14) |
| Control Variables | YES | YES | ||
| RET * Control Variables | NO | YES | ||
| D.RET * Control Variables | NO | YES | ||
| RET * D.RET * Control Variables | NO | YES | ||
| Constant and Ind. Fixed Effects | YES | YES | ||
| Observations | 2,284 | 2,284 | ||
| Adjusted R-squared | 0.225 | 0.351 | ||
| F-statistic | 19.24 | 10.82 | ||
Note(s): This table reports regression results testing whether unconditional conservatism at the start of 2020 reduced the need for conditional conservatism in firms’ 2020 income statements. The results show a significant negative association, consistent with a substitution effect between the two forms of conservatism. All variables are defined in the appendix. Industry fixed-effects, the constant term, and control variables from Table 3 are included but not reported for brevity. T-statistics of coefficient estimates are reported in parentheses and are based on robust standard errors and two-tailed tests. To facilitate the interpretation of coefficients control variables are standardized. Continuous variables are winsorized at the 1 and 99% level. *, **, and *** define significance at the 10%, 5% and 1% levels respectively
6. Conclusion
In this paper, we explore the impact of unconditional conservatism on stock market performance during the onset of the COVID-19 pandemic. Specifically, we investigate whether firms that employed conservative accounting practices experienced better stock returns and reduced investor uncertainty in the face of macroeconomic turmoil. Our results are consistent with this expectation. Additionally, we find conservatism’s effects weaker when the firm is financially healthier, the business is less volatile, the financial reports are more transparent, the business is less exposed to international markets, and the board of directors is more independent. The latter finding suggests that investors view conservatism as a corporate governance mechanism that can alleviate concerns about the firm’s prospects in times of crisis.
Our findings make several key contributions to the literature. First, we show that conservatism’s benefits to investors extend beyond its conditional form. Prior research has largely focused on conditional conservatism, likely due to the perception that it “communicates information about uncertain events, and is therefore of greater interest to researchers studying contracting and valuation issues than is unconditional conservatism” (Ruch and Taylor, 2015, p. 21). We contribute by demonstrating that unconditional conservatism can also enhance stock market outcomes by reducing valuation uncertainty during economic crises. This broadens the recognized role of conservatism in financial reporting and its value to investors.
Second, we advance understanding of how accounting shapes market dynamics during financial turbulence, a topic where research on reporting quality has yielded “mixed and inconclusive findings” (Saha, 2022). By examining conservatism’s role in sustaining abnormal returns during the pandemic, we underscore its importance for investors and the broader economy.
Finally, we contribute to the debate on conservatism’s desirability. Although regulators often view it as a reporting bias, our evidence shows it can ease investor concerns in uncertain times. This challenges the notion of conservatism as inherently harmful and highlights its potential benefits for policy, especially during crises.
In addition to its empirical findings, our study benefits from a research setting that strengthens the credibility of its inferences. The sudden and exogenous nature of the COVID-19 outbreak offers a quasi-natural experiment, reducing concerns about reverse causality or endogenous effects that can hinder the interpretation of results in archival research. Because the shock was unexpected and occurred simultaneously across firms, it allows us to more plausibly isolate the effect of pre-existing accounting practices on subsequent stock market outcomes. This strengthens the validity of our results and shows how crisis events can be used to study accounting’s role in capital markets under extreme conditions.
While our findings provide novel insights into the benefits of unconditional conservatism during the COVID-19 crisis, they should be interpreted in light of several limitations. First, although we use established empirical proxies and conduct multiple robustness checks, unobserved firm characteristics or omitted variables may still confound our results. Second, the empirical proxies we adopt, while grounded in prior literature, may fail to capture all aspects of how conservatism is implemented across firms. Third, our analysis is centered on a single macroeconomic shock which, while offering a unique research setting, may limit the generalizability of our findings to other types of crises or more stable periods. Lastly, our focus is on listed U.S. firms, and institutional differences in accounting standards, enforcement, and investor behavior may constrain the applicability of our conclusions to other countries.
These limitations suggest that further research is needed to validate and extend our findings across different settings, time periods, and forms of economic uncertainty. Future research could also investigate the specific mechanisms through which conservatism interacts with other corporate governance factors, such as managerial incentives and board characteristics, to influence firm performance during periods of turmoil. Additionally, the relationship between different forms of conservatism during crises warrants further examination, particularly to assess whether firms that deplete their hidden reserves of unconditional conservatism earlier become more reliant on conditional conservatism in response to subsequent negative shocks. Finally, future research could investigate the long-term effects of conservatism on firm performance beyond the immediate crisis period, providing a more comprehensive understanding of its role in financial reporting and market behavior. Overall, our findings highlight the critical role that accounting practices can play in reducing investor uncertainty and sustaining market stability during periods of economic turbulence.
We are grateful to the Editor-in-Chief, Prof. Tom Smith, the Associate Editor, and the two anonymous reviewers for their invaluable suggestions. We also thank the participants of the Financial Reporting and Capital Markets Workshop organized by the University of Parma in November 2024, and in particular the discussant, Prof. Tatiana Mazza, for their helpful comments.
Appendix Variable definitions
CONS_Res = Decile rank within each industry of the average value of the balance sheet reserves created by conservative accounting over the three fiscal years ending in the year t, estimated according to Penman and Zhang (2002).
CONS_Accr = Decile rank within each industry of the negative of the average non-operating accruals over the three fiscal years ending in the year t, calculated according to Givoly and Hayn (2000).
CONS_Market = Decile rank within each industry of the average end-of-year market-to-book ratio over the three fiscal years ending in the year t.
CONS = Average of the CONS_Accr, CONS_Res, and CONS_Market decile ranks.
CAR(0,45) = Cumulative abnormal returns over the (0, +45) window, where day zero is March 11, 2020.
SPREAD(0,45) = Abnormal bid-ask spreads, calculated as the average daily spread over the (0,45) window less the average spread over the (−90,−30) window, where day zero is March 11, 2020.
RETVOL(0,45) = Abnormal return volatility, calculated as the volatility of daily abnormal returns over the (0,45) window less the volatility over the (−90,−30) window, where day zero is March 11, 2020.
PriorRET = Cumulative abnormal returns over the (−90, −30) window, where day zero is March 11, 2020.
SIZE = Logarithm of total assets at the end of the fiscal year t.
LEVERAGE = Financial leverage at the end of the fiscal year t, calculated as total liabilities divided by total assets.
ROE = Return on equity for the fiscal year t−1, measured as the net income before extraordinary items scaled by the market value of equity at the beginning of the year.
PRICE = Logarithm of the stock price at the end of the fiscal year t.
FYRET = Abnormal returns cumulated over the fiscal year t.
LagFYRET = Abnormal returns cumulated over the fiscal year t−1.
ΔSALES = Year-over-year sales growth scaled by beginning total assets for the fiscal year t−1.
%IND = Quartile rank of the percentage of independent directors at the end of the fiscal year t.
#IND = Quartile rank of the number of independent directors at the end of the fiscal year t.
IND_CHAIR = Binary variable identifying if the Chairperson is an independent director at the end of the fiscal year t.
Fin.Distress = Binary variable identifying observations in the bottom quintile within the industry of the Altman’s Z-Score value averaged over the three fiscal years before the year t.
CF.Stability = Binary variable identifying observations in the bottom quintile within the industry of the operating cash flows over the prior 16 quarters.
Rep.Tranparency = Binary variable identifying observations in the bottom quintile within the industry of the volatility of the residuals of the accrual prediction model developed by Dechow and Dichev (2002), modified as per Ball and Shivakumar (2006) and McNichols (2002).
Domestic_Firm = Binary variable identifying observations in the top quintile within the industry of the ratio of domestic sales to total sales.
Less_Exposed_Industry = Binary variable identifying industries in the bottom quintile of Hassan et al.’s (2020) COVID-19 exposure measure.
RET = Buy and hold market-adjusted return over the fiscal year, as per Basu (1997).
D.RET = Binary variable identifying negative values of RET, as per Basu (1997).
SPI(t+1) = Special items reported in the t+1 income statement, scaled by the beginning market value of equity.
Notes
As stated in Section BC3.27 of Statement 8 of the FASB conceptual framework, prudence or conservatism are excluded from the fundamental characteristic of faithful representation “because including either would be inconsistent with neutrality”.
The fiscal year 2019 in 90 percent of cases, 2018 for the remaining 10 percent.
In Section 5, we create additional variables to conduct several cross-sectional tests: for example, proxies for financial distress, cash flow volatility, reporting transparency (Section 5.2), board independence (5.3), and firm exposure to international markets (5.4). Untabulated results indicate that adding these proxies as control variables in the main regression analysis leaves the coefficients of interest practically unchanged. However, limited data availability for these additional variables reduces the sample size, which in turn lowers the power of the tests and may limit the validity of the inferences.
This window captures a particularly volatile and historically unusual period in financial markets. By the time our return cumulation period begins, the broader market had already declined by approximately 20% from its February 2020 peak, reflecting the early incorporation of pandemic-related risks into stock prices. The first ten days of our window include an additional 10–15% drop, culminating in the S&P 500’s pandemic low on March 23, 2020. This was followed by a remarkably swift partial recovery, during which roughly half of the total losses were regained within just a few weeks. In robustness tests and additional analyses (e.g. Section 5.1), we report results using alternative windows and return periods within 2020.
Calculated as [(10–1) x 1.35]/28.596.
Calculated multiplying nine by each regression coefficient, respectively −0.084 and −0.029.
We are grateful to an anonymous referee for suggesting it.
Because SPIt+1 is standardized, the coefficient of 0.030 suggests that increasing CONS by one decile makes SPIt+1 less negative by an amount equal to 3 percent of the SPIt+1 sample standard deviation.

