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Noise Trading and Asset Pricing Factors∗
Shiyang Huang Yang Song Hong Xiang
May 3, 2020
Abstract
We demonstrate that a broad set of asset pricing factors (anomalies) are
significantly exposed to “noise trader risk,” and the noise trader risk is priced
in factor premia. Since mutual fund investors are ignorant of asset pricing
factors, we confirm that mutual funds’ flow-induced trades of factors are un-
informed as they have a large price impact on factor returns, followed by a
complete reversal. We then show asset pricing factors are subject to noise
trader risk in that expected variation and covariation of flow-induced noise
trading strongly forecast variance and covariance of factor returns. Impor-
tantly, we find that factor premium is higher when the flow-driven noise trader
risk is more salient.
∗We thank Bronson Argyle, Itzhak Ben-David, Philip Bond, Jaewon Choi, Lauren Cohen, Zhi Da,Larry Harris, Paul Irvine, Lukas Kremens, Hanno Lustig, Dong Lou, Stephan Siegel, Yangru Wu, YaoZeng, and seminar participants at University of Southern California, University of Washington, Universityof Colorado Boulder, Monash University, Deakin University, the University of Technology Sydney, Ts-inghua University (SEM), Renmin University of China, Shanghai Advanced Institute of Finance (SAIF),CUHK Shenzhen, Fudan Fanhai and the Central University of Economics and Finance, as well as partici-pants at the Lancaster Factor Investing Conference and the 7th Annual Melbourne Asset Pricing Meeting.Huang and Xiang are with Faculty of Business and Economics, the University of Hong Kong, and Songis with Foster School of Business, University of Washington. Email: [email protected], [email protected],and [email protected]. A previous version of this manuscript was circulated under the title “Flow-Induced Trades and Asset Pricing Factors.”
1 Introduction
Asset pricing literature has documented a broad set of stock market factors/anomalies
that explain cross-sectional variations in stock returns.1 Understanding the source and
variations of these factors has been arguably one of the main themes of asset pricing
research over the past several decades.2 In this paper, we provide a new perspective on
asset pricing factors. Specifically, we show that asset pricing factors are heavily exposed
to noise trader risk, which arises from uninformed capital allocations of mutual fund
investors. We further show that the flow-driven noise trader risk is significantly priced in
factor premia, corroborating the theory of De Long, Shleifer, Summers, and Waldmann
(1990) at the factor level.
We start by confirming our premise that the demand shift of mutual fund investors
is largely uninformed in that mutual funds’ trades induced by fund flows generate large
short-term price impact on factor returns, which reverts entirely afterward.3 Then, we
show that asset pricing factors are heavily exposed to noise trader risk as variation and
covariation of flow-induced factor trading strongly predict the variance-covariance struc-
ture among the factor returns. Importantly, when using variations of flow-induced trading
to quantify noise trader risk, we find that noise trader risk is significantly priced in factor
premium by arbitrageurs and other investors: in the time series, average premium across
factors is higher when the aggregate flow-driven noise trader risk is expected to be higher;
cross-sectionally, the return of a factor is higher when its flow-induced trading is expected
to be more correlated with the aggregate flow-driven demand.
Our analysis is motivated by recent observations that mutual fund investors are igno-
rant of asset pricing factors (Berk and van Binsbergen, 2016; Barber, Huang, and Odean,
1We use factors and anomalies interchangeably in this paper.2See, for example, Cochrane (2011), Nagel (2013), McLean and Pontiff (2016), Harvey, Liu, and Zhu
(2016), Hou, Xue, and Zhang (2018), and Harvey and Liu (2019).3Because the construction of flow-induced trading does not use contemporaneous stock returns, our
results are not subject to the critique in Wardlaw (2019).
1
2016) and respond to uninformative signals (Ben-David, Li, Rossi, and Song, 2019), when
they allocate capital among equity mutual funds. This is not surprising since the ma-
jority of mutual fund investors are households with limited information readily available
to them. For example, according to the 2011 ICI Fact Book, 93.7% of mutual fund
assets in the U.S. were held by households. Moreover, an extensive literature has also
documented that mutual fund investors exhibit behaviors that are generally considered
unsophisticated.4
As argued by Grinblatt and Titman (1989) and Pastor and Stambaugh (2002), in a
fully rational world, mutual fund investors should consider all factors that explain cross-
sectional variation in fund performance regardless of whether the factors are priced or
not, and only reward mutual fund managers with “real” alphas. In such a world, fund
flows would have little influence on factor returns. Given mutual fund investors’ actual
behaviors and their large holding of the stock market, however, we hypothesize and verify
that mutual funds’ flow-driven trades of factors are mostly uninformed, and the flow-
driven noise trading is an important state variable that is priced in factor premia.
To analyze the relationship between noise trader risk and factor premia,5 we use 70
characteristic-based stock market factors, including the Fama-French five factors and the
momentum factor. To construct a characteristic-based factor, we sort all NYSE-AMEX-
NASDAQ stocks into quintile portfolios based on the NYSE breakpoints and measure
factor returns as the spreads between the value-weighted returns of the top-quintile and
the bottom-quintile stocks.
We use a bottom-up approach to measure mutual funds’ flow-driven demand of a given
4For example, mutual fund investors prefer funds that report holdings of recent winners and lotterystocks (Solomon, Soltes, and Sosyura (2014) and Agarwal, Jiang, and Wen (2018)); invest in funds thatadvertise a lot (Jain and Wu (2000)) or appear in the media (Kaniel and Parham (2017)); prefer fundsthat recently experienced an extremely positive monthly return (Akbas and Genc (2020)); and time themarket poorly (Frazzini and Lamont (2008) and Akbas, Armstrong, Sorescu, and Subrahmanyam (2015)).Song (2019) shows that investors’ unsophisticated behaviors lead to a mismatch between managerial skilland scale of active funds.
5While our approach applies smoothly to individual stocks, in this paper we focus on asset pricingfactors as understanding source and variations of factors is one of the key topics of asset pricing research.
2
factor at a quarterly frequency. That is, we estimate mutual fund flow-induced trading
(FIT) for individual stocks in a given quarter following Lou (2012). In a nutshell, FIT
measures the magnitude of flow-driven trading by the aggregate mutual fund industry
on a particular stock over a quarter. Then, for a given factor, we use stocks’ portfolio
weights within the factor to aggregate the stock-level FIT to the factor level.6 In total,
we use 208,419 fund-quarter observations with 4,999 active equity mutual funds in the
US from 1980 to 2017.
To demonstrate that noise trader risk is priced in factor premia, we organize our
analysis in three steps. To start, we confirm that flow-induced trading of factors (FITOF)
is mostly uninformed. Specifically, we find that flow-induced trades generate large short-
term price impact on factor returns, and the flow-driven price pressure reverts entirely
afterward. For example, in a given quarter, the top 20 factors by FITOF over the same
quarter significantly outperform the bottom 20 factors by 3.75% on average (15% on an
annual basis) in terms of the Fama-French five-factor (FF5) alpha. However, when we
track their performance over a longer horizon (beyond the portfolio formation quarter),
the short-term price impact completely reverts over the following two years. This full
reversal confirms the earlier evidence that mutual fund investors are ignorant of asset
pricing factors (Berk and van Binsbergen, 2016; Barber et al., 2016; Ben-David et al.,
2019).7
In the second step of our analysis, we demonstrate that asset pricing factors are
significantly exposed to the flow-driven noise trader risk because expected variations (co-
variations) of flow-induced trading strongly forecast variations (covariations) of factor
returns. To this end, we use fund ownership and flows to estimate expected variance
and expected covariance of flow-induced trading of factors by extending the approach
6In the robustness checks in Appendix B, we also isolate fund flows that are driven by fund alphasand re-construct our measures with the “alpha-free” fund flows. We obtain similar results using the“alpha-free” fund flows.
7Moreover, we rule out reverse causality by showing that there are no return patterns when we sortfactors based on past factor returns. As a placebo test, we also show in Appendix B that the non-flow-induced trades of mutual funds do not generate factor return reversal.
3
of Greenwood and Thesmar (2011), which we refer to as “factor fragility” and “factor
co-fragility,” respectively.
We find that when mutual fund trading of a factor is expected to be more volatile,
the return volatility of this factor is indeed higher; when flow-induced trades between
factors are expected to be more correlated, factor returns covary more with each other.
For example, in the Fama-MacBeth regression of quarterly frequency, a one-standard-
deviation increase in factor co-fragility predicts an increase of 46% of a standard deviation
in factor return covariance over the next quarter, even after controlling for lagged factor
return covariance. The results remain largely unchanged when excluding the crisis periods.
Taken together, our findings indicate that the asset pricing factors are heavily exposed to
noise trader risk. In this case, the noise trader risk arises from the uninformative capital
allocation of mutual fund investors.
Our goal is to understand whether and to what extent the flow-driven noise trader
risk is “priced” by arbitrageurs and other sophisticated investors who trade these factors.8
According to De Long et al. (1990), noise trader risk should affect arbitrageurs’ willingness
to trade these factors: when noise trader risk is higher, arbitrageurs are less willing to
trade, and consequently, these factors have higher returns. In other words, arbitrageurs
and other investors shall demand higher compensation to trade these factors when the
noise trader risk is expected to be more salient.
To this end, we use fragility of the aggregate factor portfolio, that is, the expected
variation of flow-induced trading of the equal-weighted factor portfolio, to proxy for the
aggregate flow-driven noise trader risk. Based on De Long et al. (1990), the average
premium across factors should be higher when the aggregate fragility is higher. Similarly,
in the cross-sections, the required return of a factor should also be higher if flow-induced
trading of that factor is expected to be more correlated with the aggregate flow-driven
demand. Specifically, we measure this trading covariation by the co-fragility between the
8Hanson and Sunderam (2014) and McLean and Pontiff (2016), among others, show that arbitrageurs,such as hedge funds or quant funds, have widely exploited asset pricing anomalies.
4
factor and the aggregate factor portfolio.
Consistent with the intuition, aggregate factor fragility significantly and positively
forecasts future average factor premia. For example, in the time-series regression, a one-
standard-deviation increase in aggregate fragility of the 70 factors forecasts an increase
of about 60 bps in average factor premium over the next quarter. This is economically
significant as the average premium across the set of factors is about 78 bps per quarter.
The results continue to hold after controlling for the market sentiment of Baker and
Wurgler (2006), the average value spread of factors, past average factor return, and the
average factor return covariance, suggesting that the average factor co-fragility indeed
captures information beyond these predictors. Moreover, the out-of-sample (OOS) tests of
Welch and Goyal (2007) further confirm the strong predictive power of aggregate fragility
on future factor premia.
In the cross-sectional tests, we also find that the required premium of a factor is
significantly higher when its flow-driven noise trading is more correlated with the ag-
gregate flow-driven demand. On average, a factor’s expected return increases by 51 bps
per quarter with a one-standard-deviation increase in its co-fragility with the aggregate
factor portfolio. In sum, both the time-series and cross-sectional results indicate that the
flow-driven noise trader risk is priced in factor premia.
To further corroborate our claim that arbitrageurs price the flow-driven noise trader
risk, we explore hedge funds’ trading activities. Specifically, we decompose each factor
into two “sub-factors” based on hedge fund trading volume. We find that the flow-driven
risk is priced in factor premia mostly through stocks that are actually traded by hedge
funds, consistent with that arbitrageurs require compensation for bearing noise trader
risk. In contrast, other measures of sentiment (e.g. Baker and Wurgler (2006)) mostly
affect factor premia through small-cap stocks and stocks that are often not traded by
hedge funds and other institutions. This sharp difference also suggests that flow-induced
noise trading is orthgonal to existing measures of market sentiment.
5
Our paper is closely related to the literature that studies the role of non-fundamental
trading on asset prices. Examples include Shiller (1981), Lee, Shleifer, and Thaler (1991),
Campbell and Cochrane (2000), Lettau and Ludvigson (2001), and Baker and Wurgler
(2006). We complement this literature in two aspects: First, we offer the first analysis of
noise trader risk on asset pricing factors and find strong evidence that noise trader risk
is priced in factor premia. Second, unlike the prior studies that either propose measures
of market sentiment or analyze how levels of market sentiment affect stock returns, we
directly quantify noise trader risk for tested assets and find that, both in the time-series
and in the cross-sections, higher noise trader risk is associated with a higher expected
return. Moreover, while market sentiment (e.g., Baker and Wurgler, 2006)) has larger
effects on small-cap stocks, the flow-induced noise trader risk is priced mostly through
large-cap stocks, which are more likely to be traded by hedge funds and other institutional
investors.9
Another strand of literature shows that investor demand unrelated to fundamentals
can affect asset prices. For example, Coval and Stafford (2007), Frazzini and Lamont
(2008), and Lou (2012) show that mutual fund flow-induced demand shocks have a con-
siderable price impact on individual stock prices. Teo and Woo (2004) and Li (2019)
document that mutual fund flows negatively predict style-level stock returns, consistent
with the “style investing” hypothesis in Barberis and Shleifer (2003). This paper comple-
ments this literature by comprehensively analyzing the impact of uninformative demand
shocks on a large collection of asset pricing factors. Our results highlight that asset pricing
factors are heavily exposed to “noise trader risk,” which we further show is significantly
priced by arbitrageurs.
Our paper is also closely related to the recent literature that investigates the high
dimensionality of cross-sectional asset pricing models.10 We offer a new perspective on
9Stambaugh, Yu, and Yuan (2012) compare the performance of asset pricing anomalies following theperiods of high and low sentiment. Specifically, they argue that the presence of short-sale constraintdrives anomalies, particularly the short legs of anomalies, stronger following high levels of sentiment.
10Examples include Harvey, Liu, and Zhu (2016), Harvey (2017), McLean and Pontiff (2016), Kozak,
6
asset pricing factors by demonstrating the important influence of noise trading on factor
premia and the variance-covariance structure among these factors.
The rest of the paper is organized as follows. Section 2 introduces the dataset, the set of
asset pricing factors, and our measure of flow-induced trading of factors (FITOF). Section
3 confirms that flow-induced factor trades are nonfundamental. Section 4 quantifies the
influence of flow-induced trading on return volatilities and return comovements among
the factors. Section 5 shows that the flow-driven noise trader risk is significantly priced
in factor premia. Section 6 concludes. Robustness checks and supplementary results are
reported in the appendices.
2 Data and Methodology
In this section, we describe the data, the construction of the 70 asset pricing factors
(anomalies), and how we estimate mutual fund flow-induced trading of factors.
2.1 Factor Construction
We use the CRSP and Compustat datasets to construct 70 asset pricing factors. Table
C.1 shows the list of factors, which include size, book-to-market ratio, profitability, and
momentum, among many others.
Our sample stocks include all ordinary common shares (CRSP share code 10 or 11)
listed on NYSE, AMEX, and NASDAQ. To be included in our sample in a given quarter,
the stock is required to be held by at least one mutual fund with non-missing holding data
from Thomson Reuters CDA/Spectrum database and valid fund flow (calculated using
CRSP mutual fund database) in that quarter.
The universe of factors consists of 70 annually or quarterly rebalanced factors based on
Nagel, and Santosh (2019), Hou, Xue, and Zhang (2018), Kelly, Pruitt, and Su (2019), Feng, Giglio, andXiu (2019), among others.
7
firm characteristics.11 To construct a characteristic-based factor, we follow Hou, Xue, and
Zhang (2018) and use NYSE breakpoints of the characteristic to form quintile portfolios.
For annually rebalanced factors, at the June-end of each calendar year, we sort all stocks
into quintiles based on the NYSE breakpoints of sorting variables (e.g., book-to-market
ratio) at the fiscal year ending in the previous calendar year. We then track the value-
weighted portfolio returns from July to next June. For quarterly rebalanced factors that
rely on Compustat quarterly fundamentals data, we skip one quarter between the portfolio
formation date and the start of the portfolio holding period to ensure that all information
is available upon portfolio formation. Specifically, at the end of each quarter, we form
quintile portfolios based on sorting variables as of the fiscal quarter ending in the previous
calendar quarter, and hold the portfolios in the next calendar quarter. More details of
the factor construction are provided in Appendix C.
2.2 Estimate Mutual Fund Flow-Induced Trades of Factors
To measure flow-induced trades of each of the 70 factors, we first estimate flow-induced
trades of individual stocks. To this end, we merge the Thomson Reuters CDA/Spectrum
database with the CRSP Survivorship-bias-free mutual fund database. In particular, we
obtain mutual funds’ holding data from the CDA/Spectrum database. Mutual funds’
total net assets (TNA), monthly net returns (after fee), and annual expense ratios are
from the CRSP database. For mutual funds with multiple share classes, we use the sum
of TNA across all share classes as the TNA of the fund, and we take TNA-weighted
average net returns and expense ratios across all share classes. We compute mutual fund
monthly gross returns (before fee) as the sum of monthly net returns and 1/12 of the
annual expense ratio.
We focus on actively-managed equity mutual funds. Specifically, we filter out non-
11We transform several typical monthly rebalancing factors (e.g., momentum) into quarterly rebal-ancing factors to match the quarterly mutual fund holdings data. Our results do not change even afterexcluding those monthly rebalancing factors.
8
equity funds based on investment objective codes reported in the CDA/Spectrum database
and the CRSP mutual fund database.12 In addition, we require the ratio of common stock
holdings to TNA to be between 80% and 105% on average over the sample period. Finally,
we exclude fund-quarter observations with less than $1 million TNA. Our fund sample
includes 4, 999 distinct US domestic equity funds with 208, 419 fund-quarter observations
during 1980-2017.
We take two steps to construct the stock-level flow-induced trading. We first calculate
quarterly mutual fund flow, defined as the percentage change of total net assets after
adjusting for the appreciation of fund holdings (Sirri and Tufano, 1998):
Flowk,t =TNAk,t − TNAk,t−1 × (1 +Rk,t)
TNAk,t−1
,
where TNAk,t is the total net assets of fund k at the end of quarter t and Rk,t is the gross
return of fund k in quarter t.
Second, we measure quarterly aggregate mutual fund trading of an individual stock in
response to fund flows. Specifically, we follow Lou (2012) and estimate the flow-induced-
trading (FIT) measure as follows :
FITj,t =
∑k Sharesk,j,t−1 × Flowk,t × PSFk,t∑
k Sharesk,j,t−1
, (1)
where Sharesk,j,t−1 is the number of shares of stock j held by fund k at the end of quarter t−
1, Flowk,t is the percentage flow of fund k in quarter t, and PSF is the partial scaling factor.
12We mainly follow Kacperczyk, Sialm, and Zheng (2008) to screen funds in the following steps. First,we screen funds by investment objectives reported by CDA/Spectrum database. We exclude funds withInvestment Objective Codes in 1, 5, 6, or 7, in the CDA/Spectrum database. Then, we screen funds byinvestment objectives reported by CRSP mutual funds database. For funds with non-missing “Type ofSecurities Mainly Held by Fund” variable (policy variable), we remove those with policy in C&I, Bal,Bonds, Pfd, B&P, GS, MM, or TFM. We then require remaining funds to have Lipper ClassificationCode in EIEI, G, LCCE, LCGE, LCVE, MCCE, MCGE, MCVE, MLCE, MLGE, MLVE, SCCE, SCGE,SCVE, or Missing. For funds with missing Lipper Classification Code, we require them to have StrategicObjective Insight Code in AGG, GMC, GRI, GRO, ING, SCG or missing. If a fund has both missingLipper Classification Code and Strategic Objective Insight Code, we screen them through WiesenbergerFund Type Code and retain funds with objective codes in G, G-I, AGG, GCI, GRI, GRO, LTG, MCG,SCG, or Missing.
9
The scaling factor reflects how fund managers, on average, increase and liquidate their
holdings in response to capital inflows and outflows, respectively. Lou (2012) estimates
PSFk,t to be 0.970 for outflows, and 0.858 for inflows. We use the same estimates of the
partial scaling factor in our study.13 Moreover, we use FIT rather than the entire realized
trading of mutual funds because FIT only captures those trades that are driven by the
demand shifts from mutual fund investors, which are largely uninformative (Ben-David,
Li, Rossi, and Song, 2019). For robustness, in Appendix B, we re-construct FIT using
mutual fund flows that are not driven by fund alpha components, and the results remain
largely unchanged.
Based on stock-level flow-induced trading, we measure flow-induced trading of a factor
π as the value-weighted average FIT of stocks in the factor’s long leg minus the value-
weighted average FIT of stocks in the short leg. That is,
FITOFπ,t =∑j∈NπL
µπj,t−1FITj,t −∑j∈NπS
µπj,t−1FITj,t, (2)
where N πL and N π
S are the set of stocks consisting of the long-leg and short-leg of factor π
at time t, respectively, and µπj,t−1 is the weight of stock j in factor π. In short, FITOF mea-
sures the flow-induced trading of the long-leg stocks relative to the flow-induced trading
of the short-leg stocks.
[Table 1 Here]
Table 1 reports the basic statistics of our sample. As shown in Panel A, The sample
coverage steadily rises as the relative size of the mutual fund sector grows substantially
over time (from 2.62% to 20.07%). At the beginning of the sample period, our sample
covers 46.30% in terms of the number of stocks and 94.45% in terms of market capital-
ization, indicating that mutual funds tend to avoid tiny stocks, which is consistent with
Frazzini and Lamont (2008). At the end of the sample period, our sample covers 94.34%
13Our results are not sensitive to the choices of PSF.
10
in terms of the number of stocks and 99.15% in terms of market capitalization. It is also
worthy to note that because we use the value-weighted scheme to construct factor returns
or factor-level flow-induced trading, including stocks that are not held by mutual funds
(e.g., tiny stocks) in the construction of factors has little influence on our results.
As shown in Panel B of Table 1, the average quarterly return of the 70 factors is 0.78%,
with a standard deviation of 6.54%. The 25th and 75th percentiles of the stock-level FIT
are −1.95% and 3.02%, respectively. This suggests that, in response to retail investors’
demand shifts, mutual funds adjust their stock holdings relative to their existing holdings
at a scale between −1.95% and 3.02% within a quarter in the 25th to 75th percentile
range. The 25th and 75th percentiles of FITOF are −0.53% and 0.55%, respectively.
3 Flow-Induced Factor Trading is Uninformed
In this section, we justify the premise that mutual funds’ flow-induced trades of factors
are largely uninformed.14 We find that flow-induced trades significantly and positively
influence contemporaneous factor returns, followed by full reversals over longer horizons.
This return pattern confirms the prior findings that mutual fund flows are mostly unin-
formed and ignorant of systematic factors (Barber, Huang, and Odean, 2016; Ben-David,
Li, Rossi, and Song, 2019).
To examine the return pattern associated with flow-induced trading of factors (FITOF),
at each quarter-end, we sort the 70 factors into three groups based on FITOF over the
same quarter, with 20/30/20 factors in each group respectively. We then track each group
with equal-weighted factor returns in the next 12 quarters. Table 2 reports the monthly
(adjusted) returns of the three portfolios of factors sorted by FITOF.
[Table 2 Here]
14While some earlier work, such as Coval and Stafford (2007), Frazzini and Lamont (2008), and Lou(2012), shows that mutual fund flows generate price pressure on individual stocks, it is not clear whetherthe flow-induced price impact would cancel out at the factor level, as factors are constructed to bediversified long-short portfolios.
11
The first pattern to note is that FITOF generates a strong price impact on factors
over the contemporaneous quarter (Qtr 0). For example, Panel D of Table 2 reports that
the low-FITOF group earns an average monthly Fama-French five-factor (FF5) alpha of
−0.42% in the formation quarter, while the high-FITOF group earns a monthly FF5 alpha
of 0.82%. The high-minus-low return spread is associated with an average monthly FF5
alpha of 1.25% with a t-statistic of 5.93.
Second, the short-term return effect strongly reverts in the subsequent two years. For
example, when we adjust the returns for the FF5 factors (Panel D), the high-minus-low
spread is indistinguishable from zero in the first post-formation year, but it is −0.33%
per month in the second year. In the third year, there is no further reversal associated
with portfolios ranked by FITOF.
[Figure 1 Here]
To visualize the return pattern, Figure 1 plots the cumulative FF5 alpha of a long-short
portfolio that longs the high-FITOF group and shorts the low-FITOF group. As one can
see, the positive FF5 alpha of the long-short portfolio in the formation quarter almost
fully reverts by the end of the second year. This return pattern validates the premise
that flow-induced trades of factors are mostly uninformed. In addition, by exploiting
the reversal effect associated with FITOF, we find that a tradable strategy that longs
the factors with low flow-induced trades in the past eight quarters and shorts the factors
with high past FITOF generates annualized alphas of 7.3% to 11.6% depending on the
benchmark (see Table B.6).
We conduct several placebo tests in Appendix B. First, to ensure that the above return
dynamics are not driven by the mean reversion of factor returns, we conduct a similar
portfolio-sorting exercise but instead of sorting on FITOF, we use factor returns as the
sorting variable. We do not find any return patterns (see Panels A and B of Table B.3).
We also analyze the influence of mutual fund trades of factors that are not driven by fund
12
flows.15 We find that non-flow-induced trades do not experience factor return reversals
(see Panel C of Table B.3). These placebo tests highlight the uniqueness of the return
patterns associated with flow-induced factor trades.
In addition to the placebo tests, we conduct a comprehensive robustness check in
Appendix B. First, to account for time and factor fixed effects that might be correlated
with factor returns, we estimate panel regressions of factor returns on flow-induced trading
of factors. The regression exercise confirms the strong influence of fund flows on factor
returns (see Table B.2). Furthermore, we conduct the regression exercise in the first-
and second-half sample periods (1982-1998 versus 1999-2017), respectively. We find that
the effects of FITOF on factor returns are stronger in the second-half sample, which is
consistent with the rapid growth of the mutual fund industry over time.
In summary, the results in this section and in Appendix B demonstrate that the flow-
driven demand shifts are statistically significant and economically strong drivers of factor
returns. More importantly, our findings confirm the mounting evidence that mutual fund
flows are largely uninformative.
4 Factors are Exposed to Noise Trader Risk
In this section, we estimate the extent to which asset pricing factors are exposed to
noise trader risk. Specifically, we quantify noise trader risk by estimating the variance-
covariance matrix of flow-induced trading among the set of factors. We find that (i) the
expected volatility of flow-induced trading strongly forecasts future factor return volatility
and (ii) the expected covariance of flow-induced trading of factors strongly forecasts future
factor return covariance. Together with the results in Section 3 that mutual funds’ flow-
induced factor trades are uninformed, we conclude that these 70 asset pricing factors are
15The non-flow-induced trades are the difference between mutual funds’ realized trades and the flow-induced trades. To calculate mutual funds’ realized trades of factors, we first compute mutual funds’aggregate realized trades on each stock (RT) in a similar way as we calculate FIT. Then we computemutual funds’ realized trades of a factor as portfolio-weighted average RT on stocks that constitute thatfactor.
13
significantly exposed to noise trader risk, which arises from the uninformative fund flow
movements (De Long et al., 1990).
4.1 Estimate (Co)variation of Flow-Induced Factor Trading
We first describe how we estimate the variance-covariance matrix of flow-induced trad-
ing of factors. Following the spirit of Greenwood and Thesmar (2011), the expected vari-
ance of flow-induced trading of a given factor π over quarter t + 1, which we refer to as
“factor fragility,” can be estimated by
Gπt = W π
t
′Et(Ωt+1)W π
t . (3)
Here, Et(Ωt+1) is the conditional variance-covariance matrix of mutual fund flows in quar-
ter t + 1 and W πt =
(wπ1,t, . . . , w
πK,t
)′is the vector of mutual fund weights in factor π. In
particular, the weight of mutual fund k in factor π is calculated as
wπk,t =∑j
µπj,tSharesk,j,tShroutj,t
,
where µπj,t is the weight of stock j in factor π,16 and Shroutj,t is the number of shares
outstanding of stock j.
Likewise, we estimate the expected covariance of flow-induced trading of factors π1
and π2 over quarter t+ 1, referred to as “factor co-fragility,” by
Gπ1,π2t = W π1
t
′Et(Ωt+1)W π2
t . (4)
As one can see, factor fragility and factor co-fragility depend on mutual fund ownership
concentration and the expected variance-covariance matrix of mutual fund flows. To
estimate Et(Ωt+1), we calculate the variance-covariance matrix of mutual fund flows using
16For a long-leg stock, µπj,t equals its original weight in the long leg. For a short-leg stock, µπj,t is itsoriginal weight in the short leg multiplied by negative one.
14
observations in the past two years, and the results are not sensitive to this choice. We
report the detailed estimation in Appendix A.
To illustrate the role of flow variation on factor return variation, we plot the return
volatility and the square root of lagged factor fragility of the Fama-French size and value
factors in Figure 2. At first glance, there is a clear positive correlation between future
factor return volatility and volatility of flow-induced trading of factors. We formally
explore the association between factor fragility (co-fragility) and factor volatility (co-
variance) in the next subsection.
[Figure 2 Here]
4.2 Variations of Flow-Induced Trading and Return Variations
We now examine how factor fragility and factor co-fragility, our measures of expected
variance and expected covariance of flow-induced trades of factors, respectively, can fore-
cast future variance and covariance of factor returns.
To this end, we first estimate the following Fama-MacBeth regression:
σπ1,π2t+1 = α + βGπ1,π2t + τσπ1,π2t + γZπ1,π2
t + επ1,π2t+1 , (5)
where σπ1,π2t+1 is the return covariance between factors π1 and π2 in quarter t + 1 and is
estimated based on weekly factor returns, and Gπ1,π2t is the estimated co-fragility between
factors π1 and π2 in equation (4). To account for persistence in factor return covariance, we
include lagged factor return covariance. We also include a set of control variables, Zπ1,π2t ,
that comprises the factor-pairwise difference in size, book-to-market, and momentum as
in Anton and Polk (2014).17 For easy interpretation, all variables are normalized to have
17We construct factor pairwise characteristics difference as follows. First, following Anton and Polk(2014), we construct a stock-level NYSE percentile ranking of characteristics. Second, we take value-weighted NYSE percentile rankings for each of the long and short legs and compute factor-level NYSEpercentile ranking as the long-short difference. The factor pair-level difference is the absolute value ofthe difference in factor-level NYSE percentile ranking of characteristics.
15
a standard deviation of one.
[Table 3 Here]
Panel A of Table 3 reports the results. The expected covariance of flow-induced trades
between factors strongly predicts future factor return comovements. In the univariate
regression, a one-standard-deviation increase of factor co-fragility predicts an average in-
crease of 80% of a standard deviation of factor return covariance over the next quarter.
Even after controlling for lagged factor return covariance, a one-standard-deviation in-
crease of factor co-fragility still predicts an increase of 46% of a standard deviation of
factor return covariance. In columns (3) and (4), we exclude the crisis periods (from 2000
to 2001 and from 2007 to 2008), and confirm that our results are not driven by crisis.
In fact, we get even stronger effects after excluding crisis periods. These results indicate
that the covariation in flow-induced trading is an important determinant of factor return
comovement.
We then estimate the predictability of factor fragility on future factor return variation
through the following Fama-MacBeth regression:
σπt+1 = a+ b√Gπt + cσπt + ηt+1. (6)
Here, the dependent variable σπt+1 is the one-quarter-ahead factor volatility, estimated as
the standard deviation of weekly factor returns over the next quarter. The independent
variable of interest is the square root of factor fragility. We also control for past factor
return volatility.
Panel B of Table 3 reports the results. Across all specifications, we find that the square
root of factor fragility,√Gt, positively and significantly forecasts the one-quarter-ahead
factor return volatility. For example, in the univariate regression excluding the crisis
period (column 3), a one-standard-deviation increase in√Gt predicts an increase of 37%
of a standard deviation of factor volatility in the next quarter. After controlling for past
16
volatility, a one-standard-deviation increase in√Gt still leads to an average increase of
17% of a standard deviation of factor return volatility (column 4).
In summary, the results in Sections 3 and 4.2 indicate that flow-induced factor trad-
ing, while being driven by uninformative capital allocation of mutual fund investors,
significantly determines average returns, variations, and covariations among the broad
set of asset pricing factors (anomalies). In other words, these asset pricing factors are
significantly exposed to noise trader risk, which arises from the uninformative fund flow
movements. We further explore the asset pricing implications of the flow-driven risk in
the next section.
5 Is the Flow-Driven Noise Trader Risk Priced?
We have shown that mutual fund flow movements are largely uninformative, yet drive
a large portion of variations and covariations of factor returns. In this section, we explore
whether the flow-driven noise trader risk is priced in factor premia by arbitrageurs and
other investors.
Through both in-sample and out-of-sample time-series tests, we find that average
premia across factors are much higher when the aggregate flow-driven risk is expected
to be more salient. Through cross-sectional tests, we find that the required return of a
factor is higher when its flow-induced trading is expected to be more correlated with the
aggregate flow-driven risk. We also find that these “pricing” effects are mainly driven
by stocks that are traded extensively by hedge funds and driven by large-cap stocks that
institutional investors tilt towards (Gompers and Metrick, 2001). In sum, our findings
suggest that flow-driven noise trader risk is indeed priced by arbitrageurs, which confirms
the theory of De Long et al. (1990) at the factor level.
17
5.1 In-Sample Time-Series Tests
According to De Long et al. (1990), noise trader risk should affect arbitrageurs’ will-
ingness to trade factors: when noise trader risk is higher, arbitrageurs are less willing to
trade, and consequently, factors have higher returns. In this section, we show that the
average expected return across factors is indeed higher when the aggregate flow-driven
risk is higher.
Specifically, we estimate the following predictive time-series regression:
1
70
70∑i=1
rπi,t+1 = α + β Fragilityt
(1
70
70∑i=1
πi
)︸ ︷︷ ︸
Aggregate Fragility
+γ′Kt + ηt+1. (7)
Here, the dependent variable is the equal-weighted return of the 70 factors in quarter t+1.
The independent variable of interest is the fragility of the agggregate portfolio of the 70
factors, which measures the expected variation of flow-induced trading of the aggregate
factor portfolio. Kt is a vector of controls that potentially predicts future factor premium,
including the investor sentiment measure of Baker and Wurgler (2006) (BW), the average
value spread of factors (Ilmanen et al., 2019), and past average factor returns.
To set the stage, Table 4 reports the correlations between aggregate factor fragility,
Fragilityt(∑70
i=1 πi/70), and those predictors in the control Kt of equation (7). The corre-
lation between aggregate factor fragility and the BW sentiment is 0.15, and the correlation
with the average value spread is 0.30. These fairly low correlations indicate that aggregate
fragility shall capture information beyond these other predictors.
[Table 4 Here]
Table 5 presents the time-series regression of future average factor premia on aggregate
factor fragility in equation (7). For easy interpretation, all independent variables are
normalized to have a standard deviation of one. From column (1), aggregate factor
18
fragility positively and significantly forecasts future average factor premia (t = 3.22). In
terms of the economic magnitude, a one-standard-deviation increase in aggregate factor
fragility is associated with an increase of 60 bps in the average factor premia over the
next quarter. Considering that the average factor premia is about 78 bps per quarter in
our sample, the effect of noise trader risk on factor premia is economically important.
[Table 5 Here]
For comparison, columns (2)-(4) of Table 5 report the predictability of other predictors
on future factor premia. In column (2), we find that average factor return covariance also
positively forecasts future average factor returns. This result is consistent with the finding
of Pollet and Wilson (2010) that the average comovement of individual stock returns
predicts average stock returns. Columns (3) and (4) show that the BW sentiment and the
average value spread of factors are also strong predictors of future average factor premia,
consistent with Stambaugh, Yu, and Yuan (2012) and Ilmanen et al. (2019).
To confirm that the predictability of aggregate factor fragility is not driven by the
correlation with other predictors, we further conduct pairwise horse-race regressions in
columns (5) to (7) of Table 5. Column (5) shows that the predictive power of aggregate
fragility is not subsumed by average factor return covariance. In contrast, the lagged
average return covariance becomes statistically insignificant with the presence of factor
fragility. Meanwhile, aggregate fragility remains statistically significant with mild reduc-
tions in coefficient estimates when controlling for the BW sentiment or the average value
spread. Finally, we put all these predictors together into the predictive regression. Col-
umn (8) shows that only aggregate fragility remains statistically significant, while other
predictors lose the power of predicting future average factor premia.
To adjust for potential small-sample bias in the predictive regressions (Stambaugh,
1999), we also apply the bias-reduction estimation approach proposed in Amihud and
Hurvich (2004). Table 6 reports the results. Panel A shows that aggregate factor fragility
has very low auto-correlation, and Panel B shows that the innovation in the predictive
19
regression of future factor premia on aggregate fragility is almost uncorrelated with the
innovation in the AR(1) regression of aggregate fragility. Hence, the predictive regression
does not suffer from the small sample bias. Panel C confirms that the results are almost
unchanged after bias-correction.
[Table 6 Here]
We further conduct several robustness tests. First, we detrend aggregate factor
fragility to alleviate the effect of time trend on our regression analysis.18 We find that
the results are almost unchanged (columns (3)-(6) of Table 7). Second, we confirm that
the results are not specific to the particular way that we construct factors. Specifically,
we form factors with the NYSE decile breakpoints of characteristic variables and repeat
the in-sample regression.19 Under this alternative factor construction method, aggregate
fragility remains to be a statistically significant predictor of future average factor premia,
and its magnitude is even higher (columns (1) and (2) of Table 7).20
[Table 7 Here]
In summary, the predictive time-serires regressions indicate that when the aggregate
flow-driven risk is expected to be higher, other investors indeed require higher average
premia for these factors.
5.2 Out-of-sample Time-Series Tests
Welch and Goyal (2007) point out that many variables with in-sample forecasting
power cannot forecast returns out-of-sample. To further validate the role of aggregate
18In our sample period 1982Q1-2017Q4, we consider residuals from the OLS regression of aggregatefragility on the year-quarter indicator as linearly-detrended aggregate fragility. Similarly, quadratic trendis excluded through regressing aggregate fragility on both level and square terms of the year-quarterindicator.
19All factor-related variables are also re-constructed based on the decile portfolios.20 In Appendix Table B.8, we show that aggregate fragility cannot predict stock market returns and
bond market returns. In contrast, the BW sentiment index or the stock market variance can significantlypredict the long-term yield or the default yields. This indicates that the “noise trader” risk arising fromequity mutual fund flows does not capture the information contained in variables like the BW sentimentor stock market variance.
20
factor fragility in predicting future factor premia, in this section, we conduct out-of-
sample (OOS) tests as in Welch and Goyal (2007). In particular, our OOS test uses only
real-time data of a given predictor to forecast future average factor premia. Then the
OOS predictive power of the predictor is evaluated against that of the historical average
factor premia. We find that aggregate factor fragility strongly forecasts future average
factor premia in a series of OOS tests.
Specifically, we conduct the OOS test by estimating the following predictive regression
recursively:
Rt+1 = α + βAt + εt. (8)
Here, Rt+1 is the average factor return over quarter t + 1, and At refers to a specific
return predictor (e.g., aggregate factor fragility in equation (7)). Starting with an initial
in-sample estimation period, we estimate the above predictive regression and obtain the
OLS estimates (αt, βt) of (α, β). We then forecast average factor premia over the next
quarter by
Rt+1 = αt + βtAt. (9)
At each quarter, we expand the estimation window by one quarter until we reach the end
of our sample period.
To evaluate the OOS performance for a given predictor of future average factor premia,
we follow Welch and Goyal (2007) and compute the following OOS statistics:
R2OOS = 1− MSEA
MSEN
and ∆RMSE =√
MSEN −√
MSEA.
Here,
MSEA =T−1∑t=n
(Rt+1 − Rt+1
)2
, MSEN =T−1∑t=n
(Rt+1 − Rt+1
)2,
where T is the total number of quarters in our sample, n is the number of in-sample
quarters used for the first forecast, Rt+1 is the actual average factor premium, Rt+1 is the
21
forecast of average factor premium in (9), and Rt+1 is the historical mean of the average
factor premia. Intuitively, R2OOS and ∆RMSE are positive when the forecast errors based
on the predictor At, MSEA, are smaller than the forecast errors of the historical mean,
MSEN . To test the hypothesis, we also compute the MSE-F statistic by (T − h + 1) ×
((MSEN −MSEA)/MSEA) , and we compare it against the asymptotic critical values in
McCracken (2007).21
Table 8 reports the OOS performance of aggregate factor fragility and other predictors
of average factor premia used in Table 5. We choose a long enough evaluation period from
1992Q1 to 2017Q4, which starts ten years after the first quarter in our sample.22 As one
can see, aggregate factor fragility, the BW sentiment, and average value spread all have
superior OOS predictive power for future average factor premia. For instance, R2OOS
of aggregate fragility is 8.07% in the evaluation period. By comparison, average factor
return covariance fails to beat the historical mean in the OOS tests, although it has
positive in-sample predictability (Table 5).
[Table 8 Here]
Taken together, the results in Tables 5 to 8 indicate that aggregate factor fragility
positively forecasts future average factor premia both in-sample and out-of-sample. As
higher aggregate fragility implies higher noise trader risk, this evidence indicates that
arbitrageurs and other investors demand higher average premia trading these factors
when the flow-driven noise trader risk is more salient.
5.3 Explore Arbitrageurs’ Trading Activities
In this section, we further corroborate the claim that flow-driven “noise trader” risk
is priced by arbitrageurs and other sophisticated investors.
21h is the degree of overlap. Here, h = 1 for no overlap.22Hansen and Timmermann (2012) suggest that the power of forecast evaluation tests is stronger with
longer out-of-sample periods. Our choice of evaluation period ensures that the out-of-sample period islong enough relative to the initial estimation period. We also report OOS test results based on alternativechoices of evaluation periods in Table B.9, and the results are robust.
22
Specifically, we explore arbitrageurs’ trading activities. Our hypothesis is that noise
trader risk should affect factor premia only through stocks that are indeed traded by
arbitrageurs and other investors. In other words, if we construct factors using stocks
with similar factor-characteristics but different likelihood to be traded by arbitrageurs,
we should expect to find a stronger relationship between noise trader risk and factor
premia, when factors are constructed with stocks that are more likely to be traded by
arbitrageurs.
As hedge funds are typical arbitrageurs who often make use of long-short equity strate-
gies, we use hedge funds’ trading volume of a stock to gauge whether this stock is inten-
sively traded by arbitrageurs. In addition, motivated by previous literature that insti-
tutional investors trade larger stocks more than smaller stocks (Gompers and Metrick,
2001), we also construct factors using large-cap stocks and small-cap stocks separately in
a similar exercise.
Specifically, for each factor, we further sort the stocks in the long leg or the short leg
of that factor into two even groups based on their hedge fund trading volume (HF-Trade)
in the previous year.23 We then reconstruct two “sub” factors by only using stocks in
the low or the high HF-Trade group. Take the momentum factor as an example. We
construct the so-called “Low-HF-Trade” momentum factor by longing the winner stocks
in the low HF-Trade group and shorting the loser stocks that are also in the low HF-Trade
group. Similarly, the “High-HF-Trade” momentum factor is constructed by longing the
winner stocks in the high HF-Trade group and shorting the loser stocks that are also in
the high HF-Trade group.
Within the low HF-Trade factors or the high HF-Trade factors, we repeat the in-
sample and out-of-sample predictability tests of aggregate fragility on future average factor
23We calculate hedge funds’ trading volume of a stock in a given quarter using the quarterly changein hedge funds’ holdings reported in the 13F files. Then we aggregate the absolute dollar trading volumeof all hedge funds on each stock in each year. Missing hedge fund trading volume is set to zero.
23
premia. That is, similar to equation (7), we estimate
1
70
70∑i=1
rhigh HFπi,t+1 = α + βFragilityt
(1
70
70∑i=1
πhigh HFi
)+ γ′Kt + ηt+1, (10)
and
1
70
70∑i=1
rlow HFπi,t+1 = α + βFragilityt
(1
70
70∑i=1
πlow HFi
)+ γ′Kt + ηt+1, (11)
where the factors are constructed by stocks with high and low hedge fund trading volumn,
respectively.
Panel A of Table 9 reports the in-sample predictive regressions in equations (10) and
(11). We see a sharp contrast in the coefficient estimates of aggregate fragility between
the low and high HF-Trade factors (columns (1)-(4)). For example, in the univariate
regressions, the coefficient of aggregate fragility is 0.05 (t = 0.25) for low HF-Trade group,
while it is 0.59 (t = 2.74) for high HF-Trade group. This comparison beomes more striking
when we take into account the fact that the unconditional factor returns are higher for
factors constructed by stocks with low hedge fund trading volume.24
[Table 9 Here]
Panel B of Table 9 reports the out-of-sample tests in Table 8 for the low and the high
HF-Trade group of factors, respectively. Consistent with the in-sample regression results,
the out-of-sample performance of aggregate fragility is positive and significant in the high
HF-Trade group while it is slightly negative and insignificant, for the low HF-Trade group.
In a similar exercise, we also use small-cap stocks and large-cap stocks to re-form the
factors. By comparing factors formed with small and large stocks, we find the significant
relationship between aggregate fragility and average factor premia only exists among
24Average factor returns in low and high HF-Trade groups are 1.16% and 0.76% per quarter, respec-tively.
24
factors in large market capitalization group (columns (5)-(8)), consistent with Gompers
and Metrick (2001) that institutional investors tend to trade large-cap stocks.25
In short, the results in Table 9 indicate that the influence of flow-driven risk on future
factor premia is mostly through large-cap stocks and through stocks that are heavily
traded by hedge funds. By contrast, the predictability of the BW sentiment on factor
returns is mostly through stocks with high limits-to-arbitrage (e.g., stocks with low hedge
fund trading volume or small-cap stocks).
5.4 Evidence from Cross-Sectional Factor Returns
The time-series analysis so far shows that arbitrageurs require higher average factor
premia when the aggregate flow-driven noise trader risk is higher. Cross-sectionally, when
a factor’s flow-induced trading is expected to be more correlated with the aggregate flow-
driven demand, arbitrageurs should also require a higher premium for trading that factor.
We formally test this cross-sectional prediction in this section.
To measure the expected covariance of a factor’s flow-driven trading and the aggregate
noise trading, we compute the co-fragility between the factor and the aggregate factor
portfolio by
Co-Fragilityπi,t ≡ Co-Fragilityt
(πi,
1
70
70∑j=1
πj
). (12)
That is, a higher Co-Fragility implies that a factor is expected to experience more flow-
induced trading when the aggregate flow-driven demand is higher.
With this, we test the cross-sectional relationship between Co-Fragility and expected
factor returns through the Fama-MacBeth regression:
rπi,t+1 = α + βCo-Fragilityπi,t + γYπi,t + επi,t+1. (13)
25We also sort factor long/short leg into terciles by market capitalization and conduct the in-sampleand out-of-sample tests within each size tercile. Figure 4 shows that the correlation between averageco-fragility and future average factor premia monotonically increases with size tercile ranking, and suchrelation is only statistically significant in the largest size tercile.
25
Here, the dependent variable is the return of factor πi in quarter t+1, and the independent
variable of interest is the average co-fragility between factor πi and the aggregate factor
porfolio at the end of the prior quarter, as defined in equation (12). For control variables
Yπi , we include the average return covariance between factor πi and the factor portfolio
in the prior quarter (dubbed by Covariance), the past-one-quarter factor returns, the
factor value spread at the previous quarter-end, and the average FITOF in the past eight
quarters.
Table 10 reports the regression results. For easy interpretation, all independent vari-
ables are normalized to have a standard deviation of one. Across all specifications, higher
co-fragility between a factor and the aggregate factor portfolio predicts higher future
factor returns. Column (1) of Table 10 shows that a one-standard-deviation increase in
Co-Fragility is associated with an increase of 51 bps in factor return per quarter, and the
effect is statistically significant at the 1% level.26 This magnitude is also economically
significant, given that our sample average quarterly factor return is around 78 bps. Fur-
thermore, the predictive power of Co-Fragility is unchanged after including other control
variables (column (3)).
[Table 10 Here]
To alleviate the concerns of outliers, we also use the quintile ranking of Co-Fragility
in each quarter as the regressor in columns (2) and (4) of Table 10. The coefficients
remain statistically significant. To provide an easy interpretation of the economic mag-
nitude, in column (5), we only retain factors assigned to the lowest or highest quintile of
Co-Fragility each quarter and regress factor returns on a dummy variable, which indicates
the highest quintile ranking (Dummy Rank5). Column (5) shows that the coefficient of
Dummy Rank5 is 1.59 (t = 2.70), which means a long-short strategy that longs (shorts)
26In the Fama-MacBeth regression, we standardize all independent variables, except Rank andDummy Rank5, by their standard deviation in each time period. Hence, the coefficient reported hererepresents the change in monthly factor returns associated with a one-cross-sectional-standard-deviationincrease in the independent variable.
26
factors in the highest (lowest) quintile of Co-Fragility produces returns of 159 bps per
quarter.
We also show the positive return predictability of Co-Fragility through portfolio anal-
ysis. At each quarter-end, we sort factors into quintiles based on Co-Fragility in the
quarter and hold the quintile portfolios in the next quarter. As shown in Table B.10,
the CAPM alpha (FFC four-factor alpha or FF five-factor alpha) increases monotoni-
cally with Co-Fragility. For example, the monthly CAPM alpha increases from 0.02 % to
0.69% from the bottom to the top quintile. The spread of monthly CAPM alpha is 0.67%
(t = 3.73). This return pattern is also shown in Figure 3.
[Figure 3 Here]
In summary, both the time-series and the cross-sectional analyses in this section indi-
cate that the flow-driven noise trader risk on factors is an important state variable that
is priced by arbitrageurs, consistent with the theoretical predictions of De Long, Shleifer,
Summers, and Waldmann (1990).
6 Conclusion
Stock market factors (anomalies) are one of the building blocks of asset pricing re-
search. In this paper, we provide a new perspective on asset pricing factors. That is, we
demonstrate that asset pricing factors are heavily exposed to noise trader risk, where the
noise trader risk arises from uninformative demand shifts of retail mutual fund investors.
We also show that the flow-driven noise trader risk is priced in factor premia, indicating
that the predictions of De Long, Shleifer, Summers, and Waldmann (1990) also apply to
asset pricing factors.
Specifically, we take a bottom-up approach to measure mutual funds’ flow-induced
trading of asset pricing factors. We first show that the flow-induced trades have large
price impacts on contemporaneous factor returns, which fully revert afterward. This
27
evidence justifies that the flow-induced trades of factors are mostly uninformed. We then
show that the uninformative flow movements strongly forecast variation and covariation
among the asset pricing factors, indicating that factors are subject to noise trader risk.
More importantly, in both the time-series and the cross-sections, we find that arbitrageurs
and other investors require higher factor premia when the flow-driven noise trader risk is
expected to be more salient.
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33
Table 1: Summary Statistics
This table reports the summary statistics of mutual funds, stocks, and factors inour sample. Panel A reports the summary statistics of the US equity mutual funds inour study. # Funds is the number of distinct mutual funds in each period. TNA is theaverage fund total net assets (in million $). % Coverage of stock (EW) is the number ofdistinct stocks held by mutual funds in our sample, divided by the total number of CRSPstocks. % Coverage of stock (VW) is the total market capitalization of distinct stocksheld by mutual funds in our sample, divided by total market capitalization of CRSPstocks. % Market is the average percentage of the US common stocks held by the mutualfunds in our sample. Panel B reports the stock and factor characteristics. Size andbook-to-market ratio of our sample stocks are shown in NYSE percentiles. Stock-levelflow-induced trading (FIT) is defined in (1). Flow-induced trading of factor (FITOF) isdefined as the value-weighted FIT of a factor’s long-leg stocks minus that of the short-legstocks in (2). The definitions of factor-level square root of fragility (
√Fragility) and
factor pairwise co-fragility are in Section 4.1. The list of factors is in Table C.1.
Panel A: Summary statistics of mutual funds
Period # Funds TNA % Coverage of stock % Market
Median Mean EW VW
1980-1984 370 64.43 159.62 46.30 94.45 2.621985-1989 610 79.18 264.52 58.58 97.15 4.331990-1994 1,453 71.81 299.02 65.76 98.51 7.511995-1999 2,699 110.79 698.13 72.56 98.40 13.192000-2004 3,461 120.31 837.97 86.77 99.54 15.352005-2009 3,636 172.73 1,097.92 92.77 99.61 18.522010-2014 2,875 297.49 1,664.22 91.29 98.44 18.612014-2017 2,216 479.88 2,757.40 94.34 99.15 20.07
Panel B: Summary statistics of stocks and factors
Variables Mean SD Q1 Median Q3
Stock level:Size 0.3105 0.2923 0.0547 0.2143 0.5190Book-to-Market 0.4839 0.3033 0.2088 0.4801 0.7534FIT 0.0157 0.1196 −0.0195 0.0017 0.0302Factor level:Quarterly Ret 0.0078 0.0654 −0.0270 0.0048 0.0388FITOF 0.0003 0.0177 −0.0053 0.0001 0.0055SD of Daily Ret 0.0060 0.0041 0.0037 0.0049 0.0068√
Fragility 0.0012 0.0014 0.0005 0.0008 0.0015Factor-pair level:Cov of Daily Ret (10−6) 0.4683 6.0821 −0.5313 0.1116 0.8814Co-Fragility (10−6) 0.0097 0.7377 −0.0377 0.0021 0.0514
34
Tab
le2:
Retu
rnpatt
ern
of
flow
-in
duce
dtr
adin
gof
fact
ors
This
table
rep
orts
the
per
form
ance
offa
ctor
por
tfol
ios
sort
edby
flow
-induce
dtr
adin
gof
fact
or(F
ITO
F).
FIT
OF
isth
ep
ortf
olio
-wei
ghte
dflow
-induce
dtr
adin
g(F
IT)
ofa
fact
or’s
long-
leg
stock
sm
inus
that
ofth
esh
ort-
leg
stock
s(s
eeeq
uat
ion
(2))
.In
each
quar
tert,
we
sort
the
70fa
ctor
sin
toth
ree
grou
ps
bas
edon
thei
rF
ITO
Fin
quar
tert,
wit
h20
/30/
20fa
ctor
sin
the
low
/mid
/hig
hgr
oup,
resp
ecti
vely
.E
ach
fact
oris
give
neq
ual
wei
ght
inth
ere
spec
tive
por
tfol
io.
The
por
tfol
ios
are
rebal
ance
dev
ery
quar
ter
and
hel
dfo
rth
ree
year
s.Q
uar
ter
0is
the
por
tfol
iofo
rmat
ion
quar
ter.
We
trac
kth
em
onth
lyca
lendar
-tim
ere
turn
sof
fact
orp
ortf
olio
sfr
omquar
ter
0to
quar
ter
12.
To
dea
lw
ith
over
lappin
gp
ortf
olio
sin
each
hol
din
gm
onth
,w
efo
llow
Jeg
adee
shan
dT
itm
an(1
993)
toca
lcula
teeq
ual
-wei
ghte
dre
turn
sof
por
tfol
ios
form
edin
diff
eren
tquar
ters
.P
anel
sA
toD
rep
ort
aver
age
mon
thly
raw
retu
rns,
mon
thly
CA
PM
alpha,
mon
thly
Fam
a-F
rench
-Car
har
tfo
ur-
fact
oral
pha,
and
mon
thly
Fam
a-F
rench
five
-fac
tor
alpha
duri
ng
1980
-201
7,re
spec
tive
ly.
The
retu
rns
and
alphas
are
rep
orte
din
per
cent.
Thet-
stat
isti
csin
par
enth
eses
are
com
pute
dbas
edon
stan
dar
der
rors
wit
hN
ewey
-Wes
tco
rrec
tion
for
twel
vela
gs.
Portfolio
Qtr
0Qtr
1-4
Qtr
5-8
Qtr
9-12
Portfolio
Qtr
0Qtr
1-4
Qtr
5-8
Qtr
9-12
PanelA:M
onth
lyRaw
Retu
rnPanelB:M
onth
lyCAPM
Alpha
Low
−0.2
20.4
1∗∗
∗0.4
9∗∗
∗0.
27∗∗
∗L
ow−
0.10
0.55∗∗
∗0.6
3∗∗
∗0.
42∗∗
∗
(−1.
61)
(3.4
2)(3
.86)
(2.7
0)
(−0.
70)
(4.1
2)
(4.8
7)
(4.7
9)
Mid
0.2
6∗∗
∗0.
25∗∗
∗0.2
7∗∗
∗0.
25∗∗
∗M
id0.
35∗∗
∗0.
34∗∗
∗0.3
5∗∗
∗0.
35∗∗
∗
(4.5
5)(4
.48)
(4.5
7)
(4.1
8)
(6.6
5)
(6.4
9)
(6.3
0)
(5.7
1)
Hig
h0.7
6∗∗
∗0.
16∗∗
0.0
50.2
4∗∗
∗H
igh
0.85∗∗
∗0.
23∗∗
∗0.1
3∗∗
0.33∗∗
∗
(8.0
4)(2
.21)
(0.8
1)
(3.4
1)
(8.5
3)
(3.5
0)
(2.0
0)
(3.7
3)
H−
L0.
97∗∗
∗−
0.25
∗−
0.44∗
∗∗−
0.03
H−
L0.
95∗∗
∗−
0.32∗
−0.
50∗
∗∗−
0.08
(5.1
8)(−
1.74
)(−
2.87)
(−0.
27)
(4.5
1)
(−1.
90)
(−3.
04)
(−0.
66)
PanelC:M
onth
lyFFC4Alpha
PanelD:M
onth
lyFF5Alpha
Low
−0.1
9∗
0.3
6∗∗
∗0.3
9∗∗
∗0.
26∗∗
∗L
ow−
0.42∗
∗∗0.2
0∗∗
∗0.3
6∗∗
∗0.
23∗∗
∗
(−1.
72)
(4.9
8)(4
.13)
(2.8
5)
(−4.
43)
(2.8
3)
(3.3
8)
(2.6
8)
Mid
0.2
1∗∗
∗0.
21∗∗
∗0.2
2∗∗
∗0.
22∗∗
∗M
id0.
18∗∗
∗0.
18∗∗
∗0.1
8∗∗
∗0.
18∗∗
∗
(4.5
4)(5
.30)
(5.0
7)
(4.7
8)
(3.1
6)
(4.1
8)
(4.3
5)
(3.4
2)
Hig
h0.6
5∗∗
∗0.
100.0
50.1
8∗∗
∗H
igh
0.82∗∗
∗0.
19∗∗
∗0.0
30.
12∗
(5.6
6)(1
.57)
(0.8
7)
(3.2
6)
(6.5
5)
(2.5
9)
(0.5
1)
(1.7
5)
H−
L0.
84∗∗
∗−
0.26
∗∗−
0.33∗∗
−0.
08
H−
L1.
25∗∗
∗−
0.01
−0.
33∗
∗−
0.12
(3.9
9)(−
2.40
)(−
2.48)
(−0.
65)
(5.9
3)
(−0.
05)
(−2.
25)
(−0.
89)
35
Table 3: Predicting factor return covariance and volatility
Panel A reports the Fama-MacBeth regressions of one-quarter-ahead factor pair-wise return covariance (σπ1,π2t+1 ) on the factor pairwise co-fragility (Gπ1,π2
t ). σπ1,π2t+1 is thecovariance of weekly returns between factors π1 and π2 in quarter t+1. Gπ1,π2
t is theco-fragility between factors π1 and π2 measured at the end of quarter t, following thedefinition in equation (4). The control variables include the one-quarter lagged pairwisereturn covariance and pairwise differences in size, book-to-market, and momentum. PanelB reports the Fama-MacBeth regressions of one-quarter-ahead factor return volatility(σt+1) on the square root of factor fragility (
√Gt). σt+1 is the standard deviation of
weekly factor returns in quarter t+1 and√Gt is the square root of factor fragility, which
is defined in equation (3). In columns (1)-(2), we report the estimates of the full sampleperiod. In columns (3)-(4), we exclude observations in the crisis period (from 2000 to2001 and from 2007 to 2008). For easy interpretation, all variables are standardizedto have unit variance. *, **, *** indicate significance at the 10%, 5%, and 1% level,respectively.
Panel A: Predict pairwise factor return covariance
DepVar: σπ1,π2t+1 (1) (2) (3) (4)
Full Sample Exclude Crisis PeriodGπ1,π2t 0.80∗∗∗ 0.46∗∗∗ 1.06∗∗∗ 0.60∗∗∗
(4.96) (5.64) (4.55) (5.51)σπ1,π2t 0.53∗∗∗ 0.53∗∗∗
(16.18) (15.53)Size Diff −0.03∗∗∗ −0.04∗∗∗
(−7.26) (−7.14)BM Diff 0.01 0.02∗∗
(1.50) (2.27)MOM Diff −0.00 −0.01
(−0.40) (−1.38)
No. Obs. 347,760 347,760 309,120 309,120Adj. R2 0.10 0.34 0.10 0.34
Panel B: Predict factor return volatility
DepVar: σt+1 (1) (2) (3) (4)
Full Sample Exclude Crisis Period√Gt 0.27∗∗∗ 0.12∗∗∗ 0.37∗∗∗ 0.17∗∗∗
(7.49) (5.10) (9.26) (5.77)σt 0.63∗∗∗ 0.62∗∗∗
(25.54) (26.04)
No. Obs. 10,080 10,080 8,960 8,960Adj. R2 0.07 0.41 0.07 0.39
36
Table 4: Correlation between aggregate fragility and other sentiment measures
This table reports the pairwise correlations among aggregate fragility and severalsentiment measures. Aggregate fragility is the fragility calculated based on the equal-weighted portfolio of 70 factors in each quarter. Avg Covariance is the average pairwisereturn covariance of the factors in a given quarter. BW Sentiment is the investorsentiment index from Baker and Wurgler (2006) in the last month of a given quarter.Avg Value Spread is the average value spread of the factors at the end of a givenquarter. The value spread of a factor is computed as the log difference between theportfolio-weighted book-to-market ratio of the long-leg and the short-leg. Avg FactorRet is the equal-weighted average quarterly returns of the factors in a given quarter.
(1) (2) (3) (4) (5)
(1) Aggregate Fragility 1.00
(2) Avg Covariance 0.23 1.00
(3) BW Sentiment 0.15 0.78 1.00
(4) Avg Value Spread 0.30 0.41 0.25 1.00
(5) Avg Factor Ret 0.27 0.33 0.16 0.75 1.00
37
Table 5: Aggregate fragility and future average factor premia
This table reports the estimation results from the predictive regressions of averagefactor premia on aggregate fragility. The dependent variable is the equal-weightedaverage quarterly returns (in percent) of the 70 factors in quarter t+ 1. The independentvariables include the fragility calculated on the equal-weighted portfolio of the 70 factorsin quarter t (Aggregate Fragility), the average pairwise daily return covariance of thefactors in quarter t (Avg Covariance), the investor sentiment index from Baker andWurgler (2006) in the last month of quarter t (BW Sentiment), the average value spreadof the factors at the end of quarter t, and average quarterly returns of the factors inquarter t. The sample period is from 1982Q1 to 2017Q4. For easy interpretation,all independent variables are standardized to have unit variance. The t-statistics inparentheses are computed based on standard errors with Newey-West correction of fourlags. *, **, *** indicate significance at the 10%, 5%, and 1% level, respectively.
(1) (2) (3) (4) (5) (6) (7) (8)
Aggregate Fragility 0.60∗∗∗ 0.56∗∗∗ 0.45∗∗∗ 0.42∗∗∗ 0.43∗∗
(3.22) (3.11) (2.71) (3.00) (2.41)Avg Covariance 0.32∗ 0.19 −0.10
(1.83) (1.43) (−0.44)BW Sentiment 0.70∗∗∗ 0.58∗∗∗ 0.25
(3.60) (2.93) (0.82)Avg Value Spread 0.74∗∗∗ 0.62∗∗∗ 0.42
(3.02) (2.90) (1.03)Avg Factor Ret 0.15
(0.65)
No. Obs. 144 144 144 144 144 144 144 144Adj. R2 0.08 0.02 0.11 0.12 0.08 0.15 0.15 0.15
38
Table 6: Bias adjustment of the predictive regression in Table 5
This table reports the analyses of small-sample bias in the predictive regression ofaverage factor premia on aggregate fragility. The dependent variable is the equal-weighted average quarterly returns (in percent) of the 70 factors in a given quarter t+ 1,and the independent variable is the fragility calculated on the equal-weighted portfolioof the 70 factors in quarter t (Aggregate Fragility). The sample period is from 1982Q1to 2017Q4. Panel A reports the OLS estimates from the following two equations: AvgFactor Rett+1 = a + b × Aggregate Fragilityt + ut+1 and Aggregate Fragilityt+1 = c + d× Aggregate Fragilityt + vt+1. Panel B reports correlations or standard deviations(shown in brackets) of the innovations in the two regressions above. Panel C reports thecoefficient estimates and t-statistics (shown in parentheses) of the predictive regressionbased on the bias-reduction estimation approach in Amihud and Hurvich (2004). *, **,*** indicate significance at the 10%, 5%, and 1% level, respectively.
Panel A: Original OLS estimates
a b c d
0.58∗∗∗ 0.60∗∗∗ 0.45∗∗∗ −0.07
(3.19) (3.63) (5.00) (−0.86)
Panel B: Correlation [SD]
u v
u [1.98] 0.06
v [1.00]
Panel C: Bias-adjusted estimates
ac bc
0.58∗∗∗ 0.60∗∗∗
(3.19) (3.62)
39
Table 7: Robustness checks for predicting average factor premia
This table reports several robustness checks for the predictive regressions of aver-age factor premia on factor co-fragility. First, we re-construct factors through formingvalue-weighted long-short portfolios with NYSE decile breakpoints of characteristicvariables. All factor-level variables are defined following Table 5 but are constructedusing NYSE decile long-short portfolios. The regression results are reported in columns(1) and (2). We also detrend the key independent variable Aggregate Fragility in theregressions. Specifically, in the sample period of 1982Q1-2017Q4, we regress AggregateFragility on a year-quarter time indicator and use the residuals as linear-detrended AvgCo-Fragility. Similarly, we regress Aggregate Fragility on a year-quarter time indicatortogether with its square term and use the residuals as quadratic-detrended AggregateFragility. Regression results with the detrended Aggregate Fragility are reported incolumns (3)-(6). All independent variables are standardized to have unit variance. Thet-statistics in parentheses are computed based on standard errors with Newey-Westcorrection of four lags. *, **, *** indicate significance at the 10%, 5%, and 1% level,respectively.
(1) (2) (3) (4) (5) (6)NYSE Decile Portfolio Detrend-Linear Detrend-Quadratic
Aggregate Fragility 0.74∗∗∗ 0.51∗∗ 0.61∗∗∗ 0.43∗∗ 0.60∗∗∗ 0.43∗∗
(3.22) (2.39) (3.25) (2.39) (3.36) (2.43)Avg Covariance −0.01 −0.09 −0.09
(−0.03) (−0.43) (−0.39)BW Sentiment 0.55∗ 0.25 0.23
(1.71) (0.82) (0.74)Avg Value Spread 0.30 0.41 0.43
(0.75) (1.02) (1.07)Avg Factor Ret 0.05 0.14 0.15
(0.23) (0.64) (0.65)
No. Obs. 144 144 144 144 144 144Adj. R2 0.08 0.15 0.08 0.15 0.08 0.15
40
Table 8: Out-of-sample tests of forecasting average factor premia
This table reports the statistics of the out-of-sample forecast errors for averagefactor premia at a quarterly frequency. We calculate the out-of-sample test statistics,R2
OOS and ∆RMSE, following Welch and Goyal (2007) to compare the predictiveregression forecast against the unconditional mean forecast. A star next to the estimatesof R2
OOS is based on the critical values of the MSE-F statistic given by McCracken (2007).The MSE-F statistic tests the equivalence of MSE of the unconditional mean forecastand the conditional forecast. The definition of each predictor is the same as in Table5. All numbers are in percent. *, **, *** indicate significance at the 10%, 5%, and 1%level, respectively.
Predictor R2OOS ∆RMSE MSE F
Aggregate Fragility 8.07∗∗∗ 0.09 12.55
Avg Covariance −16.11 −0.18 −19.84
BW Sentiment 9.47∗∗∗ 0.11 14.96
Avg Value Spread 8.03∗∗∗ 0.09 12.48
Avg Factor Ret 0.11 0.00 0.15
41
Tab
le9:
Pre
dic
ting
avera
ge
fact
or
pre
mia
for
fact
ors
form
ed
wit
hsu
bse
tsof
stock
s
This
table
rep
orts
the
resu
lts
from
the
pre
dic
tive
regr
essi
ons
ofav
erag
efa
ctor
pre
mia
inT
able
s5
and
8,w
her
efa
ctor
sar
eco
nst
ruct
edusi
ng
stock
sw
ith
diff
eren
thed
gefu
nd
trad
ing
volu
mes
orm
arke
tca
p.
Inco
lum
ns
(1)-
(4),
we
sort
the
stock
sin
the
long
orsh
ort
leg
ofa
fact
orin
totw
oev
engr
oups
bas
edon
thei
rhed
gefu
nd
trad
ing
volu
me
inth
epre
vio
us
year
.F
orea
chfa
ctor
,w
ere
-con
stru
ctva
lue-
wei
ghte
dlo
ng-
shor
tp
ortf
olio
susi
ng
stock
sin
the
low
and
hig
hhed
ge-f
und-t
radin
g-vo
lum
egr
oup,
resp
ecti
vely
.Sim
ilar
ly,
inco
lum
ns
(5)-
(8),
we
sort
the
long-
leg
orsh
ort-
leg
stock
sin
tosm
all-
cap
and
larg
e-ca
pgr
oups
bas
edon
thei
rm
arke
tca
pit
aliz
atio
nat
the
pre
vio
us
year
-end
and
re-c
onst
ruct
fact
ors
usi
ng
smal
l-ca
pan
dla
rge-
cap
stock
s,re
spec
tive
ly.
All
fact
or-l
evel
and
fact
or-p
air
leve
lva
riab
les
are
re-c
onst
ruct
edfo
llow
ing
the
defi
nit
ions
inT
able
5.F
orea
syin
terp
reta
tion
,al
lin
dep
enden
tva
riab
les
are
stan
dar
diz
edto
hav
eunit
vari
ance
.T
het-
stat
isti
csin
par
enth
eses
are
com
pute
dbas
edon
stan
dar
der
rors
wit
hN
ewey
-Wes
tco
rrec
tion
offo
ur
lags
.*,
**,
***
indic
ate
sign
ifica
nce
atth
e10
%,
5%,
and
1%le
vel,
resp
ecti
vely
.
PanelA:In
-sample
Regre
ssions
(1)
(2)
(3)
(4)
(5)
(6)
(7)
(8)
Low
HF
Tra
de
Hig
hH
FT
rad
eS
mal
lM
ktc
apL
arg
eM
ktc
ap
Agg
regate
Fra
gil
ity
0.0
5-0
.03
0.59∗∗∗
0.42∗∗
−0.
16−
0.21∗
0.58∗∗∗
0.4
2∗∗
(0.2
5)(-
0.19
)(2
.74)
(2.1
4)(−
1.23
)(−
1.94)
(2.6
2)
(2.0
9)
Avg
Cov
aria
nce
−0.0
4−
0.07
−0.
39∗
−0.0
8(−
0.4
4)(−
0.29
)(−
1.83)
(−0.3
6)B
WS
enti
men
t0.4
4∗∗
0.20
0.6
8∗∗∗
0.24
(2.5
5)(0
.67)
(3.0
0)(0
.79)
Avg
Val
ue
Sp
read
0.3
4∗0.
440.
50∗
0.42
(1.8
7)(1
.07)
(1.7
0)(1
.05)
Avg
Fact
orR
et−
0.0
00.
13−
0.26
0.1
4(−
0.0
0)(0
.60)
(−0.
97)
(0.6
2)
No.
Ob
s.144
144
144
144
144
144
144
144
Ad
j.R
2−
0.01
0.08
0.07
0.14
0.00
0.16
0.07
0.1
4
PanelB:Out-of-sa
mple
Tests
(1)
(2)
(3)
(4)
(1)
(2)
(3)
(4)
Low
HF
Tra
de
Hig
hH
FT
rad
eS
mal
lM
ktc
apL
arg
eM
ktc
ap
R2 O
OS
∆R
MS
ER
2 OO
S∆
RM
SE
R2 O
OS
∆R
MS
ER
2 OO
S∆
RM
SE
Agg
regate
Fra
gil
ity−
1.19
−0.0
17.2
5∗∗∗
0.08
−0.
300.0
07.
27∗∗∗
0.09
42
Table 10: Cross-sectional predictive regressions of factor returns
This table reports the results of the cross-sectional Fama-MacBeth regressions.The dependent variable is the return (in percent) of a factor in quarter t+1. Co-Fragilityis the co-fragility between a given factor and the equal-weighted portfolio of 70 factorsin the prior quarter. Rank is the quintile ranking of Co-Fragility in a given quarter.Dummy Rank5 is a dummy that equals one for factors in the highest Co-Fragilityquintile and zero otherwise. Covariance is the daily return covariance between a givenfactor and equal-weighted portfolio of the 70 factors in the prior quarter. Other controlsinclude factor returns in quarter t, the value spread of a given factor at the end of theprior quarter, and the average FITOF over the past eight quarters. We standardize theindependent variables by their cross-sectional standard deviations each quarter, exceptfor Rank and Rank5. Columns (1) to (4) report the regression results in the full sample.Columns (5) to (6) report the regression results among factors in the lowest and highestCo-Fragility quintile each quarter. t-statistics in parentheses are computed based onstandard errors with Newey-West correction of 12 lags. *, **, *** indicate significanceat the 10%, 5%, and 1% level, respectively.
(1) (3) (2) (4) (5) (6)
Full Sample Extreme Quintiles
Co-Fragility 0.51∗∗∗ 0.30∗∗
(2.57) (2.22)Rank 0.33∗∗ 0.18∗∗
(2.42) (2.09)Dummy Rank5 1.59∗∗∗ 0.99∗∗
(2.70) (2.42)
Covariance 0.18 0.18 0.06(0.89) (0.83) (0.26)
Past one-quarter return 0.24 0.24 0.27(1.41) (1.47) (1.50)
Value Spread 0.03 0.06 0.03(0.08) (0.28) (0.20)
Past eight-quarter FITOF −0.27∗ −0.27∗ −0.39∗
(−1.89) (−1.86) (−1.91)
Adj. R2 0.12 0.11 0.44 0.43 0.19 0.49
43
PortfolioFormation
-1 0 1 2 3 4 5 6 7 8
Event Quarter
0%
1%
2%
3%
4%
5%
6%C
um
ula
tive
FF
5 A
lph
aFITOF Hedge Portfolio
Figure 1: Cumulative FF5 alpha of the FITOF-hedge portfolio. This figure plotsthe cumulative FF5 alpha of the long-short portfolio ranked by FITOF. FITOF measuresthe mutual fund flow-induced trading for a given factor (see equation (2)). At eachquarter-end, the 70 factors are sorted into three groups based on FITOF in ascendingorder (20/30/20 factors are assigned into the low/mid/high group, respectively). Eachfactor is given equal weight in the portfolios, and the portfolios are held for three years.The long-short portfolio goes long in the high FITOF group and short in the low FITOFgroup. Quarter 0 is the portfolio formation quarter.
44
1982Q1 1990Q1 2000Q1 2010Q1 2017Q4
0.0025
0.0050
0.0075
0.0100
0.0125
0.0150
SM
B V
ola
tilit
y
0.000
0.002
0.004
0.006
SM
B F
rag
ility
SMB FragilitySMB Volatility
1982Q1 1990Q1 2000Q1 2010Q1 2017Q4
0.005
0.010
0.015
0.020
HM
L V
ola
tilit
y
0.000
0.001
0.002
0.003
0.004
0.005
HM
L F
rag
ility
HML FragilityHML Volatility
Figure 2: Factor volatility and square root of lagged factor fragility. This figureplots the return volatility and the square root of lagged factor fragility of the Fama-Frenchsize and value factors. Factor return volatility is measured as the standard deviation ofweekly factor returns over a given quarter, and factor fragility is defined in equation (3).
45
Q1 Q2 Q3 Q4 Q5 Q5-Q1
Quintile and Hedge Portfolios
0.00%
0.25%
0.50%
0.75%
1.00%M
on
thly
CA
PM
Alp
ha
Figure 3: Monthly CAPM alpha of portfolios sorted on factor-level Avg Co-Fragility. Each quarter-end, we sort the 70 factors into quintiles based on their averageco-fragility with the equal-weighted portfolio of the 70 factors (Co-Fragility) in the quar-ter, and we hold the portfolios in the next quarter. This figure shows the average monthlyCAPM alpha of the quintile portfolios (leftmost five bars) and the “quintile five minusquintile one” long-short portfolio (rightmost bar) in the holding period of January 1982to December 2017. Limit lines around the bars show the 95% confidence intervals forestimates of monthly CAPM alpha.
46
1 2 3
Size Terciles
-0.05
0.10
0.25
0.40
0.55
Co
effi
cien
t E
stim
ates
-0.3
0.6
1.5
2.4
3.3
T-s
tati
stic
T-statisticCoefficient Estimates
1 2 3
Size Terciles
-3.0%
-1.5%
0.0%
1.5%
3.0%
4.5%
6.0%
7.5%
OO
S R
-Sq
uar
ed
-0.03%
-0.02%
0.00%
0.02%
0.03%
0.05%
0.06%
0.08%
Del
ta R
MS
E
Delta RMSEOOS R-Squared
Figure 4: Predicting average factor premia for factors formed by stocks withdifferent market cap. We sort the stocks in the long or short leg of a factor into tercilesbased on their market capitalization at previous year-end (Size tercile 1 to 3 from smallestto largest) and re-construct factors using stocks in each of the size terciles separately. Wethen conduct the in-sample and out-of-sample tests across the size terciles (see Table 9for example). The top panel shows the coefficient estimate of Aggregate Fragility in theunivariate regression of predicting one-quarter-ahead average factor premia across sizeterciles. The bottom panel shows OOS test statistics (R2
OOS and ∆RMSE) across sizeterciles.
47
Appendix
A Derivation of Factor fragility and factor co-fragility
In this section, we describe how we derive the measures of factor fragility (co-fragility).
Similar to Greenwood and Thesmar (2011) (GT), we assume the following relationship
between mutual fund flow-induced trading and return of stock j:
rj,t = αj + λj
∑k Sharesk,j,t−1fk,tPSFk,t∑
k Sharesk,j,t−1
+ εj,t. (14)
Here, rj,t is the return of stock j in quarter t, Sharesk,j,t−1 is the number of shares of
stocks j held by fund k at the end of quarter t− 1, fk,t is the percentage flow of fund k in
quarter t, and PSF is the partial scaling factor as in (1). αj and λj are two parameters.
In our implementation, we assume that λj = λ∑
k Sharesk,j,t−1/Shroutj,t−1, where λ is
the unconditional price impact factor and Shroutj,t−1 is shares outstanding of stock j at
the end of quarter t− 1. The residual term, εj,t, has a conditional mean of zero and may
capture other sources of variation of returns (e.g., news about fundamentals).
Factors are value-weighted portfolios of stocks and thus the return of factor π can be
expressed as:
rπ,t =∑j
µπj,t−1rj,t, (15)
where µπj,t−1 is the weight of stock j in factor π in quarter t.27 Combining (14) and (15),
we get
rπ,t =∑j
µπj,t−1αj + λ
(∑k
wπk,t−1fk,tPSFk,t
)+∑j
µπj,t−1εj,t, (16)
where wπk,t−1 =∑
j µπj,t−1Sharesk,j,t−1/Shroutj,t−1 can be regarded as the weight of mutual
fund k in factor π in quarter t.
27For a long-leg stock, µπj,t equals its original weight in the long leg. For a short-leg stock, µπj,t is itsoriginal weight in the short leg multiplied by negative one.
48
Based on equation (16), the conditional variance and covariance of rπ,t+1 at the end
of quarter t are
Vart(rπ,t+1) = λ2W πt
′Et(Ωt+1)W π
t + Vart
(∑j
µπj,tεj,t+1
)(17)
and
Covt(rπ1,t+1, rπ2,t+1) = λ2W π1t
′Et(Ωt+1)W π2
t + Covt
(∑j
µπ1j,tεj,t+1,∑j
µπ2j,tεj,t+1
), (18)
respectively. Here, Et(Ωt+1) is the conditional variance-covariance matrix of mutual fund
flows in quarter t + 1 and W πt =
(wπ1,t, . . . , w
πK,t
)is the vector of mutual fund weights in
factor π.
Similar to GT, we define “factor fragility” of factor π in quarter t as
Gπt = W π
t
′Et(Ωt+1)W π
t . (19)
Likewise, we define co-fragility between factor π1 and factor π2 to be
Gπ1,π2t = W π1
t
′Et(Ωt+1)W π2
t . (20)
To estimate Et(Ωt+1), we calculate the variance-covariance matrix of mutual fund flows
using observations in the most recent eight quarters (including quarter t). The summary
statistics of fragility and co-fragility are reported in Table 1.
B Additional Results
This section provides supplementary information for our main results.
Table B.1 reports the quarterly transition matrix for the factor quintile portfolios
sorted on FITOF. The probability that a factor stays in the same quintile over two
49
consecutive quarters is 31% to 50%. This suggests that FITOF is not highly persistent.
Table B.2 shows the relation between FITOF and factor returns in Panel and Fama-
MacBeth regressions. After accounting for factor and time fixed effects, we still find a
strong positive relation between factor return and contemporaneous FITOF and a sizable
negative relation between factor return and average FITOF in past five- to eight-quarter.
This result confirms our findings in Table 2. Furthermore, we run the regressions of
factor returns on contemporaneous or past FITOF in the first- and second-half sample
periods separately. We find that the price effect of FITOF is stronger in the second-half
sample period (1999-2017). This is consistent with the dramatic growth of mutual fund
industries, which is also documented in the summary statistics (see Table 1).
Table B.3 examines the return pattern of factor portfolios sorted on factor returns or
mutual funds’ non-FIT trade. The portfolio analysis procedure is the same as that in
Table 2, and the only difference is the sorting variable. When sorting on factor returns,
we do not find reversals in the three years after portfolio formation. This ensures that
the reversal pattern associated with FITOF is not driven by the mean-reversion of factor
returns. We also sort on mutual funds’ non-FIT trade. We take a bottom-up approach
to compute mutual funds’ non-FIT trade of factors. First, at stock-level, we calculate
mutual funds’ non-FIT trade on a stock as mutual funds’ aggregate realized trade scaled
by the total number of shares held by mutual funds (dubbed by RT) minus flow-induced
trading (see 1). Second, we compute mutual funds’ non-FIT trade of a factor as the
portfolio-weighted average non-FIT trade of stocks that constitute the factor. We find no
statistically significant return patterns associated with mutual funds’ non-flow-induced
trades on factors. This highlights the unique non-fundamental feature of mutual funds’
flow-induced trading on factors.
In Table B.4 and Table B.5, we isolate fund flows that are driven by fund alphas and
re-construct our key measures (FITOF, factor fragility, and factor co-fragility) with the
alpha-isolated fund flows. In these robustness tests, the return patterns of FITOF and the
50
predictive power of factor fragility (co-fragility) on factor return volatility (covariance)
are close to those in the main tables.
Table B.6 shows the performance of a trading strategy that longs factors with low
past FITOF and shorts factors with high past FITOF. We compute past FITOF for a
given factor as follows. For each stock in the long-short portfolio of the factor in a holding
period that belongs to quarter t, we compute its average FIT during Qtr t − 5 to Qtr
t− 8. Then we compute portfolio-weighted average past FIT for each factor based on its
portfolio composition in quarter t as past FITOF. At the beginning of each quarter, we
sort the 70 factors into quintiles by their past FITOF and long (short) the lowest (highest)
past FITOF quintile for one quarter. Each factor is given equal weight in the portfolio,
and the portfolios are rebalanced quarterly. During April 1982 to December 2017, such a
trading strategy generates an average monthly raw returns of 0.79% (t-statistic = 3.05)
and a CAPM alpha of 0.97% (t-statistic = 3.53).
We conduct several robustness tests for Table 5. First, we decompose the aggregate
fragility of the 70-factor portfolio into average pair-wise co-fragility of the 70 factors (Avg
Co-Fragility) and average fragility of the 70 factors (Avg Fragility). We then re-conduct
the regression analysis as in Table 5 but with the two new variables. Table B.7 shows that
only the Avg Co-Fragility can positively and significantly predict future average factor
premia. Second, in Table B.8, we conduct a placebo test using average co-fragility to
predict stock market returns and bond market returns. We find no results in placebo
tests.
Table B.9 considers alternative evaluation periods for the out-of-sample test and finds
that the OOS performance is robust. It is worthy to note that the OOS performance of
aggregate fragility is stronger in the later periods.
Finally, we conduct two robustness checks for Table 10. First, we use portfolio sorting
analysis to show the positive return predictability of Co-Fragility. At each quarter-end,
we sort factors into quintiles based on Co-Fragility in the quarter and hold the quintile
51
portfolios in the next quarter. Table B.10 shows that factor returns (alphas) increase
monotonically with factor-level average co-fragility. Second, we decompose Co-Fragility
into a factor’s average co-fragility with the rest of the factors and the factor’s own fragility,
and we re-conduct the regression analysis in Table 10 using these two variables as key
independent variables seperately. Table B.11 shows that only a factor’s average co-fragility
with the rest of the factors can positively and significant forecasts its future returns.
Table B.1: Transition matrix of FITOF quintile portfolios
In each quarter, we sort the 70 factors into quintiles based on the FITOF in thatcurrent quarter. This table reports the quarter-to-quarter transition likelihood for theFITOF quintile ranking.
Rank Qtr t+ 1 →Rank Qtr t ↓ 1 2 3 4 5
1 0.49 0.22 0.12 0.08 0.08
2 0.22 0.33 0.22 0.13 0.10
3 0.11 0.23 0.31 0.23 0.11
4 0.08 0.13 0.22 0.35 0.20
5 0.09 0.09 0.12 0.20 0.50
52
Table B.2: Regression analysis of factor return on FITOF
This table reports the regression of factor return on contemporaneous and pastFITOF. The dependent variable is the monthly factor return (in percent) in a givenmonth. Panel A reports the regressions of monthly factor returns on the flow-inducedtrading of factor (FITOF) in the contemporaneous quarter. FITOF is the value-weightedflow-induced trading (FIT) of a factor’s long-leg stocks minus that of the short-legstocks. Panel B reports the regressions of monthly factor returns on past FITOF. In agiven quarter, past FITOF refers to the average FITOF in the period of qtr t− 5 to qtrt − 8. Columns (1)-(2) report the regression results based on the full sample period ofApr 1982 to Dec 2017. Columns (3)-(6) report the regression results in two sub-periods:1982-1999 (first-half) and 1999-2017 (second-half). Regression method “Panel” refersto panel regression where time and factor fixed effects are included, and t-statistics arecomputed based on standard errors double clustered by factor and time. The regressionmethod “FM” refers to the Fama-MacBeth regression. *, **, *** indicate significance atthe 10%, 5%, and 1% level, respectively.
Panel A: Factor return and contemporaneous FITOF(1) (2) (3) (4) (5) (6)
Full Sample First Second First Second
Contemporaneous FITOF 0.24∗∗∗ 0.44∗∗∗ 0.15∗∗∗ 0.53∗∗ 0.27∗∗∗ 0.59∗∗∗
(3.29) (3.97) (3.14) (2.48) (3.33) (3.01)
Factor FE Yes Yes YesYear-Month FE Yes Yes YesRegression Method Panel FM Panel Panel FM FM
No. Obs. 30,030 30,030 14,070 15,960 14,070 15,960Adj. R2 0.09 0.17 0.07 0.11 0.15 0.09
Panel B: Factor return and past FITOF(1) (2) (3) (4) (5) (6)
Full Sample First Second First Second
Past FITOF −0.20∗∗∗ −0.35∗∗∗ −0.15∗∗∗ −0.28∗∗ −0.29∗∗∗ −0.41∗∗∗
(−3.29) (−4.03) (−2.65) (−2.13) (−3.28) (−2.81)
Factor FE Yes Yes YesYear-Month FE Yes Yes YesRegression Method Panel FM Panel Panel FM FM
No. Obs. 30,030 30,030 14,070 15,960 14,070 15,960Adj. R2 0.08 0.10 0.06 0.09 0.19 0.08
53
Table B.3: Return pattern of portfolios sorted by factor returns or non-FITtrades. This table reports the performance of factor portfolios sorted by factor returnor non-FIT trades. We conduct the following portfolio analysis with different sortingvariables across Panels A to C: At the end of each quarter t, we sort the 70 factors intothree groups based on a given sorting variable, with 20/30/20 factors in the low/mid/highgroup respectively. Each factor is given equal weight in the respective portfolio. Theportfolios are rebalanced every quarter and held for three years. Qtr 0 is the portfolioformation quarter. We track the monthly calendar-time returns of factor portfolios fromQtr 1 to Qtr 12. We deal with overlapping portfolios in each holding month followingJegadeesh and Titman (1993). Monthly FF5 alphas (%) are reported. The t-statisticsin parentheses are computed based on standard errors with Newey-West correction fortwelve lags. In Panel A, the sorting variable is the factor return in Qtr 0. In Panel B, thesorting variable is the cumulative factor return from Qtr −3 to Qtr 0. In Panel C, thesorting variable is mutual funds’ non-FIT trade of factors. At stock-level, we calculatemutual funds’ non-FIT trade on a stock as mutual funds’ aggregate realized trade scaledby the total number of shares held by mutual funds (dubbed by RT) minus flow-inducedtrading (see equation 1). Then we compute mutual funds’ non-FIT trade of a factor asthe portfolio-weighted average non-FIT trades of stocks that constitute the factor.
Portfolio Qtr 1-4 Qtr 5-8 Qtr 9-12
Panel A: Sort on current-quarter ret
Low0.10∗∗ 0.18∗∗∗ 0.21∗∗∗
(2.51) (2.72) (3.37)
Mid0.18∗∗∗ 0.19∗∗∗ 0.19∗∗∗
(3.82) (4.54) (3.80)
High0.30∗∗∗ 0.21∗∗∗ 0.13∗∗∗
(2.79) (3.67) (2.79)
H−L0.20 0.03 −0.08
(1.44) (0.42) (−1.08)
Panel B: Sort on past four-quarter ret
Low0.09 0.24∗∗∗ 0.21∗∗
(1.52) (2.81) (2.55)
Mid0.18∗∗∗ 0.21∗∗∗ 0.17∗∗∗
(4.17) (4.64) (3.53)
High0.31∗∗∗ 0.12∗ 0.16∗∗∗
(2.73) (1.80) (2.71)
H−L0.23 −0.11 −0.06
(1.45) (−0.98) (−0.49)
Panel C: Sort on mutual funds’ non-FIT trades
Low0.25∗∗∗ 0.20∗∗∗ 0.16∗∗
(2.78) (2.70) (2.20)
Mid0.17∗∗∗ 0.21∗∗∗ 0.19∗∗∗
(4.58) (3.86) (3.36)
High0.16∗∗∗ 0.15∗∗∗ 0.19∗∗∗
(3.75) (4.10) (5.54)
H−L−0.09 −0.05 0.03
(−0.90) (−0.60) (0.42)
54
Tab
leB
.4:
Retu
rnpatt
ern
offlow
-in
duce
dtr
adin
goffa
ctors
:excl
ude
fund
flow
com
pon
ents
dri
ven
by
fun
dalp
has
To
isol
ate
fund
flow
sth
atar
edri
ven
by
fund
alpha
com
pon
ents
,w
ees
tim
ate
afu
nd-q
uar
ter
pan
elre
gres
sion
wit
hti
me
fixed
effec
ts:
The
dep
enden
tva
riab
leis
fund
flow
inquar
tert,
and
the
indep
enden
tva
riab
les
incl
ude
FF
C4
alpha
ofth
efu
nd,
esti
mat
edusi
ng
obse
rvat
ions
inth
e24
mon
ths
pri
orto
quar
tert,
and
four
lags
offu
nd
flow
from
quar
tert−
4to
quar
ter
t−
1.W
ith
the
alpha-
excl
uded
fund
flow
s,w
ere
-con
stru
ctF
ITO
Fan
dco
nduct
the
sam
ple
exer
cise
asin
Tab
le2.
Pan
elA
toP
anel
Dre
por
tsm
onth
lyra
wre
turn
s,m
onth
lyC
AP
Mal
pha,
mon
thly
Fam
a-F
rench
-Car
har
tfo
ur-
fact
oral
pha,
and
mon
thly
Fam
a-F
rench
five
-fac
tor
alpha,
resp
ecti
vely
.T
he
retu
rns
and
alphas
are
rep
orte
din
per
cent.
Thet-
stat
isti
csin
par
enth
eses
are
com
pute
dbas
edon
stan
dar
der
rors
wit
hN
ewey
-Wes
tco
rrec
tion
for
twel
vela
gs.
Portfolio
Qtr
0Qtr
1-4
Qtr
5-8
Qtr
9-12
Portfolio
Qtr
0Qtr
1-4
Qtr
5-8
Qtr
9-12
PanelA:M
onth
lyRaw
Retu
rnPanelB:M
onth
lyCAPM
Alpha
Low
−0.
28∗∗
0.41
∗∗∗
0.49∗∗
∗0.
28∗∗
∗L
ow−
0.18
0.53∗∗
∗0.
61∗∗
∗0.
41∗∗
∗
(−2.
38)
(3.3
8)(4
.01)
(2.9
7)
(−1.
54)
(3.9
3)
(4.7
6)
(4.7
2)
Mid
0.29
∗∗∗
0.26
∗∗∗
0.26∗∗
∗0.
26∗∗
∗M
id0.
39∗∗
∗0.
35∗∗
∗0.
36∗∗
∗0.
36∗∗
∗
(4.3
7)(4
.60)
(4.3
3)
(4.2
0)
(6.1
0)
(6.8
0)
(6.2
1)
(5.7
2)
Hig
h0.
77∗∗
∗0.
15∗∗
0.0
60.2
2∗∗
∗H
igh
0.87∗∗
∗0.
23∗∗
∗0.1
4∗∗
0.32∗∗
∗
(7.7
5)(2
.10)
(0.8
9)
(2.9
9)
(8.5
9)
(3.4
3)
(2.2
3)
(3.6
6)
H−
L1.
05∗∗
∗−
0.26
∗−
0.43∗
∗∗−
0.06
H−
L1.
05∗∗
∗−
0.31∗
−0.
47∗
∗∗−
0.09
(5.8
0)(−
1.73
)(−
2.89)
(−0.
57)
(5.4
9)
(−1.
75)
(−2.
83)
(−0.
70)
PanelC:M
onth
lyFFC4Alpha
PanelD:M
onth
lyFF5Alpha
Low
−0.
22∗∗
0.37
∗∗∗
0.39∗∗
∗0.
27∗∗
∗L
ow−
0.43∗∗
∗0.
21∗∗
∗0.
35∗∗
∗0.
23∗∗
∗
(−2.
14)
(4.9
1)(4
.32)
(2.8
5)
(−4.
13)
(3.1
2)
(3.6
2)
(2.7
8)
Mid
0.23
∗∗∗
0.22
∗∗∗
0.22∗∗
∗0.
23∗∗
∗M
id0.
20∗∗
∗0.
19∗∗
∗0.
18∗∗
∗0.
19∗∗
∗
(4.8
2)(5
.33)
(4.5
5)
(5.0
4)
(3.2
1)
(3.9
8)
(3.8
4)
(3.5
4)
Hig
h0.
65∗∗
∗0.
090.0
60.1
6∗∗
∗H
igh
0.80∗∗
∗0.
18∗∗
0.0
40.1
0(5
.85)
(1.3
5)(0
.99)
(3.0
4)
(6.1
3)
(2.2
9)
(0.6
3)
(1.6
0)
H−
L0.
87**
*−
0.28
∗∗−
0.33∗
∗−
0.11
H−
L1.2
2***
−0.
03
−0.
31∗
∗−
0.13
(4.3
1)(−
2.41
)(−
2.50)
(−0.
85)
(5.4
5)
(−0.
29)
(−2.
28)
(−1.
08)
55
Table B.5: Predicting factor return covariance and volatility: exclude fund flowcomponents driven by fund alphas. We exclude fund flow components driven by fundalphas as in Table B.4. With the alpha-excluded fund flows, we re-construct factor fragilityand co-fragility and repeat the analyses in Table 3. Panel A reports the Fama-MacBethregressions of one-quarter-ahead factor pairwise return covariance (σπ1,π2t+1 ) on the factorpairwise co-fragility (Gπ1,π2
t ). Panel B reports the Fama-MacBeth regressions of one-quarter-ahead factor return volatility (σt+1) on the factor fragility (
√Gt). In columns
(1)-(2), we report Fama-MacBeth estimates in the full sample period from 1981Q1 to2017Q4. In columns (3)-(4), we exclude observations in the crisis period (years 2000,2001, 2007, and 2008). For easy interpretation, all variables are standardized to have unitvariance. *, **, *** indicate significance at the 10%, 5%, and 1% level, respectively.
Panel A: Predict pairwise factor return covariance
DepVar:σπ1,π2t+1 (1) (2) (3) (4)
Full Sample Exclude Crisis PeriodGπ1,π2t 0.41∗∗∗ 0.24∗∗∗ 0.52∗∗∗ 0.30∗∗∗
(8.77) (8.27) (7.80) (7.47)σπ1,π2t 0.52∗∗∗ 0.52∗∗∗
(15.75) (15.16)Size Diff −0.03∗∗∗ −0.04∗∗∗
(−5.56) (−6.44)BM Diff 0.01 0.01∗∗
(0.87) (1.99)MOM Diff −0.00 −0.01
(−0.04) (−1.00)
No. Obs. 347,760 347,760 309,120 309,120Adj. R2 0.13 0.35 0.12 0.34
Panel B: Predict factor return volatility
DepVar: σt+1 (1) (2) (3) (4)
Full Sample Exclude Crisis Period√Gt 0.19∗∗∗ 0.07∗∗∗ 0.28∗∗∗ 0.11∗∗∗
(10.13) (5.70) (11.20) (6.34)σt 0.63∗∗∗ 0.62∗∗∗
(25.44) (26.59)
No. Obs. 10,080 10,080 8,960 8,960Adj. R2 0.07 0.41 0.08 0.39
56
Table B.6: Trading strategy based on flow-induced trades of factors.
This table reports the performance of factor portfolios sorted by past FITOF. Wecompute past FITOF for a given factor in quarter t as follows. For each factor compo-nent stock in a holding period that belongs to quarter t, we compute its average FITduring Qtr t − 5 to Qtr t − 8. Then we compute portfolio-weighted average past FITfor each factor based on its portfolio composition in quarter t as past FITOF. At thebeginning of each quarter, we sort the 70 factors into quintiles by their past FITOFand hold the equally-weighted portfolio for one quarter. This table reports the monthlyrisk-adjusted returns for each factor portfolio in the holding period of April 1982 toDecember 2017. t-statistics are computed based on standard errors with Newey-Westcorrection for twelve lags.
Portfolio Excess CAPM FFC4 FF5
1 (L)0.63∗∗∗ 0.81∗∗∗ 0.53∗∗∗ 0.48∗∗∗
(3.66) (4.58) (3.81) (3.07)
20.46∗∗∗ 0.61∗∗∗ 0.38∗∗∗ 0.33∗∗∗
(3.83) (5.19) (4.09) (3.66)
30.28∗∗∗ 0.38∗∗∗ 0.19∗∗∗ 0.17∗∗
(3.85) (4.89) (3.87) (2.04)
40.14∗∗∗ 0.22∗∗∗ 0.14∗∗ 0.12∗
(2.59) (4.14) (2.52) (1.74)
5 (H)−0.16 −0.16 −0.12 −0.13
(−1.51) (−1.43) (−1.15) (−1.07)
1− 50.79∗∗∗ 0.97∗∗∗ 0.64∗∗∗ 0.61∗∗
(3.05) (3.53) (2.82) (2.24)
57
Table B.7: Predict average factor premia by Avg Co-Fragility and Avg Fragility
This table reports the estimation results from the predictive regressions of averagefactor premia on the two components of Aggregate Fragility: Avg Co-Fragility and AvgFragility. The dependent variable is the equal-weighted average quarterly returns (inpercent) of the 70 factors in quarter t + 1. In Panel A, the key independent variableis the average pairwise co-fragility of the 70 factors in quarter t (Avg Co-Fragility). InPanel B, the key independent variable is the average fragility of the 70 factors in quartert (Avg Fragility). Control variables and sample period are the same as Table 5. Foreasy interpretation, all independent variables are standardized to have unit variance.The t-statistics in parentheses are computed based on standard errors with Newey-Westcorrection of four lags. *, **, *** indicate significance at the 10%, 5%, and 1% level,respectively.
Panel A: Avg Co-Fragility as Predictor
(1) (2) (3) (4) (5)
Avg Co-Fragility 0.54∗∗∗ 0.50∗∗∗ 0.40∗∗∗ 0.37∗∗∗ 0.38∗∗
(3.14) (3.21) (2.65) (2.89) (2.47)Avg Covariance 0.23 −0.07
(1.60) (−0.34)BW Sentiment 0.61∗∗∗ 0.27
(2.95) (0.88)Avg Value Spread 0.64∗∗∗ 0.42
(2.82) (1.02)Avg Factor Ret 0.13
(0.60)
No. Obs. 144 144 144 144 144Adj. R2 0.07 0.08 0.15 0.16 0.17
Panel B: Avg Fragility as Predictor
(1) (2) (3) (4) (5)
Avg Fragility 0.16 0.14 0.11 0.11 0.10(0.52) (0.47) (0.42) (0.47) (0.41)
Avg Covariance 0.31∗ −0.02(1.83) (-0.09)
BW Sentiment 0.69∗∗∗ 0.31(3.75) (1.05)
Avg Value Spread 0.73∗∗∗ 0.49(3.16) (1.09)
Avg Factor Ret 0.07(0.32)
Observations 144 144 144 144 144Adj. R2 0.00 0.01 0.11 0.12 0.11
58
Table B.8: Placebo test: Predicting other returns
This table reports the predictive regressions of one-quarter-ahead quarterly S&P500excess return, long-term yield, and T-bill rate on aggregate fragility. Definitions of thedependent variables follow Welch and Goyal (2007). *, **, *** indicate significance atthe 10%, 5%, and 1% level, respectively.
(1) (2) (3)DepVar: S&P 500 Long-term yield T-bill rate
Aggregate Fragility −1.09 −0.00 −0.00(−1.61) (−1.56) (−0.92)
Avg Covariance −1.56∗ −0.01∗∗∗ −0.01∗∗∗
(−1.89) (−5.84) (−5.38)BW Sentiment −0.22 0.02∗∗∗ 0.02∗∗∗
(−0.30) (10.92) (9.64)Stock market variance 0.52 0.01∗∗∗ 0.00∗
(0.68) (2.93) (1.95)
No. Obs. 144 144 144Adj. R2 0.05 0.46 0.40
Table B.9: OOS tests of forecasting average factor premia with differentevaluation periods
This table reports the OOS performance of average co-fragility in the forecasts ofone-quarter-ahead average factor premia. We report OOS performance in three differentevaluation periods, as indicated in the first column. All numbers are in percent. *, **,*** indicate significance at the 10%, 5%, and 1% level, respectively.
OOS Period R2OOS ∆RMSE MSE F
1987Q1-2017Q4 (5 years from sample start) 7.42∗∗∗ 0.08 11.47
1997Q1-2017Q4 (15 years from sample start) 8.31∗∗∗ 0.11 12.96
2000Q1-2017Q4 8.81∗∗∗ 0.12 13.82
59
Table B.10: Performance of factor portfolios sorted on Co-Fragility
This table reports the performance of factor portfolios sorted by past Co-Fragility.The definition of Co-Fragility follows Table 10. At the beginning of each quarter, wesort the 70 factors into quintiles by their Co-Fragility in previous quarters and hold theportfolios for one quarter. Each factor is given equal weight in the portfolio. This tablereports the average monthly raw returns, CAPM alpha, Fama-French-Carhart four-factoralpha, and Fama-French five-factor alpha for each factor portfolio in the holding periodof January 1982 to December 2017. t-statistics are computed based on standard errorswith Newey-West correction for twelve lags.
Portfolio Excess CAPM FFC4 FF5
1 0.01 0.02 -0.04 0.00(0.15) (0.27) (−0.50) (−0.03)
2 0.24∗∗∗ 0.30∗∗∗ 0.16∗∗∗ 0.20∗∗∗
(3.65) (4.30) (3.00) (2.70)3 0.26∗∗∗ 0.38∗∗∗ 0.21∗∗∗ 0.18∗∗∗
(3.48) (5.94) (3.42) (3.14)4 0.32∗∗∗ 0.46∗∗∗ 0.28∗∗∗ 0.19∗∗∗
(3.76) (6.02) (4.34) (2.84)5 0.52∗∗∗ 0.69∗∗∗ 0.52∗∗∗ 0.39∗∗∗
(4.31) (6.10) (5.10) (3.43)
5− 1 0.51∗∗∗ 0.67∗∗∗ 0.56∗∗∗ 0.39∗∗
(2.74) (3.73) (3.24) (2.01)
60
Table B.11: Cross-sectional predictive regressions of factor returns
This table reports the results of a similar Fama-MacBeth regression as in Table10 with different independent variables. In Panel A, the key independent variable,
Co-Fragility, is the average pairwise co-fragility between a given factor and the rest of
the 70 factors in the prior quarter. Rank is the quintile ranking of Co-Fragility in agiven quarter. Dummy Rank5 is a dummy that equals one for factors in the highest
Co-Fragility quintile and zero otherwise. In Panel B, the key independent variable,Fragility, is the fragility of a given factor in the prior quarter. Rank and Dummy Rank5are defined following Panel A but based on factor fragility in the quarter. Controlvariables are the same as those in Table 10. We standardize the independent variablesby their cross-sectional standard deviations for each cross-section, except for Rank andRank5. Columns (1) to (4) report the regression results in the full sample. Columns(5) to (6) report the regression results among factors in the lowest and highest quintileeach quarter. t-statistics in parentheses are computed based on standard errors withNewey-West correction of 12 lags. *, **, *** indicate significance at the 10%, 5%, and1% level, respectively.
Panel A: Average co-fragility with other factors
(1) (2) (3) (4) (5) (6)
Full Sample Extreme Quintiles
Co-Fragility 0.48*** 0.30**(2.63) (2.30)
Rank 0.33** 0.18**(2.48) (2.17)
Rank5 1.56*** 0.87**(2.76) (2.15)
Controls No No Yes Yes No YesAdj. R2 0.11 0.44 0.11 0.43 0.19 0.49
Panel B: Fragility as regressor
(1) (2) (3) (4) (5) (6)
Full Sample Extreme Quintiles
Fragility 0.09 0.06(1.16) (1.17)
Rank 0.03 0.03(0.31) (0.69)
Rank5 0.06 0.06(0.27) (0.33)
Controls No No Yes Yes No YesAdj. R2 0.01 0.01 0.40 0.40 0.02 0.43
61
C List of factors
Table C.1 lists the 70 factors studied in our paper and reports their average monthly
raw returns, CAPM alphas, and Fama-French five factor alphas during January 1980-
December 2017. We compute the sorting variables for the 70 factors (anomalies) follow-
ing Hou, Xue, and Zhang (2018), Linnainmaa and Roberts (2018), and Arnott, Clements,
Kalesnik, and Linnainmaa (2019). Since our study requires the factor long-short port-
folio to be rebalanced quarterly or annually to match with the quarterly mutual fund
holdings data, we convert several typically monthly rebalanced factors into quarterly re-
balanced ones. These factors include the 52-week high, industry momentum, intermediate
momentum, long-term reversals, maximum daily returns, momentum, customer momen-
tum, geographic momentum, industry lead-lag, segment momentum, residual momentum,
Frazzini-Pedersen beta, and idiosyncratic volatility. For these variables, we use the latest
possible sorting variables at each quarter-end to form portfolios and hold the portfolios
in the next quarter.
62
Table C.1: List of factors
Factor Raw Ret CAPM FF5 Factor Raw Ret CAPM FF5
52-week high 0.51 1.01 0.50 Industry momentum 0.44 0.26 0.36Abnormal capital investment 0.18 0.12 0.25 Intermediate momentum 0.47 0.45 0.62Accruals 0.21 0.33 0.08 Investment growth 0.19 0.30 0.02Advertising expense 0.39 0.46 −0.01 Investment-to-assets 0.13 0.12 −0.10Altman’s Z-score 0.07 −0.08 0.25 Investment-to-capital 0.08 0.40 −0.28Amihud illiquidity 0.30 0.22 0.18 Long-term reversals 0.25 0.26 0.03Analyst earnings forecast Revision 0.26 0.30 0.24 M/B and accruals 0.34 0.43 0.19Asset Growth 0.25 0.39 −0.15 Maximum daily return 0.30 0.93 0.09Book-to-june-end-market 0.20 0.28 −0.10 Momentum 0.18 0.28 0.26Book-to-market 0.19 0.28 −0.08 Net operating assets 0.33 0.31 0.27Capital turnover 0.19 0.14 0.23 Net payout yield 0.34 0.68 0.01Cash-based profitability 0.18 0.29 0.52 Number of earnings increase 0.23 0.91 0.48Cashflow-to-price 0.24 0.46 −0.01 Ohlson’s O-score 0.16 0.32 0.31Change in asset turnover 0.14 0.17 0.01 One-year share issuance 0.37 0.57 0.05Change in long-term NOA 0.29 0.35 0.00 Operating cash flow-to-price 0.31 0.46 −0.03Customer momentum 0.55 0.66 0.56 Operating leverage 0.34 0.43 0.19Debt issuance 0.14 0.17 0.01 Operating profitability 0.35 0.60 0.17Discretionary accruals 0.27 0.20 0.42 Organizational capital-to-book 0.31 0.47 0.17Distress risk 0.61 0.89 0.68 Percent accruals 0.22 0.31 0.09Earnings forecast to Price 0.37 0.61 0.09 Piotroski’s F-score 0.06 0.10 0.06Earnings persistence 0.39 0.53 0.46 Profit margin -0.05 0.16 −0.10Earnings timeliness 0.06 −0.06 0.07 QMJ profitability 0.40 0.49 0.29Earnings-to-price 0.32 0.59 0.01 R&D expense 0.44 0.26 0.36Enterprise multiple 0.25 0.25 0.41 Real estate ratio 0.30 0.22 0.42Firm age −0.01 −0.32 0.28 Residual momentum 0.56 0.67 0.55Five-year share issuance 0.40 0.54 0.21 Return on assets 0.35 0.58 0.27Frazzini-Pedersen beta 0.23 0.91 0.48 Return on equity 0.35 0.56 0.15Geographic momentum 0.32 0.37 0.37 Sales growth −0.14 −0.35 0.17Gross profitability 0.18 0.24 0.32 Sales-minus-inventory growth 0.15 0.15 0.09Growth in Inventory 0.38 0.48 0.21 Sales-to-price 0.35 0.33 −0.18Growth score 0.11 0.27 0.31 Segment momentum 0.28 0.35 0.29Idiosyncratic volatility 0.36 0.88 0.22 Size 0.17 0.05 0.04Industry adjusted CAPX growth 0.18 0.29 0.06 Sustainable growth 0.28 0.42 0.04Industry concentration 0.26 0.16 0.54 Tax expense change 0.23 0.16 0.24Industry lead-lag 0.32 0.37 0.37 Total external financing 0.25 0.53 0.04
Average 0.26 0.37 0.23
63