Institutional Trading, Trading Volume, and Spread
ABSTRACT The effect of institutional trading on the bid-ask spread is of interest to regulators and market makers. It is often (casually) argued that greater institutional participation results in increased volatility in the market. On the other hand, some argue that greater liquidity trading by institutions should reduce spread. There has been no direct empirical evidence or theoretical knowledge to suggest a relation between institutional trading and spread. In this paper, we present a clearer picture of the nature and effect of institutional trading on spreads. We show that institutional trading is not always information driven, part of it is liquidity trading in nature. Information induced trading increases the adverse selection component. However, large volume (liquidity) trading reduces the order processing costs. Moreover, institutional buys have differential information from sells. Buys reduce spreads, while sells increase the adverse selection component. The effective spread impounds the differential nature of their trading.
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ABSTRACT: Although single-stock futures (SSFs) are useful multi-purpose stock derivatives, they have not received much attention in developed markets. We analyze SSFs in the Indian market to understand their contribution in price leadership. The findings indicate that trades in the stock market contribute more to price discovery than trades in the SSF market (72% and 28%, respectively), while quotes in the SSF market are more price innovative than quotes in the stock market (39% and 61%, respectively). Our analysis suggests that while stock and SSF trade returns have predictive ability for each other, in the case of quotes, only SSF quotes have predictive ability for stock and SSF returns.Global Finance Journal 01/2009;
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ABSTRACT: The aim of this paper is to examine changes in information asymmetry around takeover an-nouncements. Empirical studies generally find evidence of a significant reaction of stock prices, trading volumes or bid-ask spreads before the formal announcement date, but they do not consider separately the different components of the spreads, nor do they take into account the implications of the market microstructure on the reaction of traders to takeovers announcements. Using high fre-quency data, we investigate empirically the behavior of the components of 70 target firms bid-ask spreads around takeover announcements on the Paris Bourse between January 1995 and Decem-ber 2000. We apply the Lin, Sanger and Booth (1995) method to estimate these components, i.e. adverse selection costs, order processing costs and an order persistence component. Our results con-firm previous findings on French markets as we do not find evidence of the presence of informed traders before announcement dates. After the announcement, results suggest a decrease in informa-tion asymmetry, as adverse selection costs felt by around 60 %. Order processing costs, representing the limit order trader's gross profit, also decline by 54 %, probably because of the higher competition between liquidity suppliers, driving to a global reduction of the bid-ask spread. We conclude that takeovers announcements allow the release of information about target firm value, but that traders do not react before the Bloomberg press release.05/2002;
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ABSTRACT: Using a novel information asymmetry index based on measures of adverse selection developed by the market microstructure literature, we test whether information asymmetry is an important determinant of capital structure decisions, as suggested by the pecking order theory. Our index relies exclusively on measures of the market's assessment of adverse selection risk rather than on ex ante firm characteristics. We find that information asymmetry does affect the capital structure decisions of U.S. firms over the sample period 1973--2002. Our findings are robust to controlling for conventional leverage factors (size, tangibility, Q ratio, profitability), the sources of firms' financing needs, and such firm attributes as stock return volatility, stock turnover, and intensity of insider trading. For example, we estimate that on average, for every dollar of financing deficit to cover, firms in the highest adverse selection decile issue 30 cents of debt more than firms in the lowest decile. Overall, this evidence explains why the pecking order theory is only partially successful in explaining all of firms' capital structure decisions. It also suggests that the theory finds support when its basic assumptions hold in the data, as should reasonably be expected of any theory. The Author 2008. Published by Oxford University Press on behalf of The Society for Financial Studies. All rights reserved. For Permissions, please email: email@example.com, Oxford University Press.Review of Financial Studies 08/2009; 22(8):3211-3243. · 4.75 Impact Factor
Institutional Trading, Trading Volume, and Spread
Malay K. Dey
University of Massachusetts at Amherst
Northern Illinois University
University of Minnesota
[This Draft: March 2001]
We are grateful to NYSE for providing us with the TORQ data and to Ben Branch,
Shanta Hegde, and Robert Miller for their comments and numerous helpful suggestions.
All comments are welcome.
Please address correspondence to: Malay K. Dey, Dept. of Finance, Northern Illinois
University, Dekalb, IL 60115, Tel. (815) 753-6395, Email: firstname.lastname@example.org, OR B.
Radhakrishna, Department of Accounting, Carlson School of Management, University of
Minnesota, 321, 19th Avenue S., #3-122, Minneapolis, MN 55455, Tel. (612) 624-0302,
Fax (612) 626-1335, Email: email@example.com.
Institutional trading, Trading Volume, and Spread
Besides its academic interest, the effect of institutional trading on the bid-ask
spread is of interest to regulators and market makers. It is often (casually) argued that
greater institutional participation results in increased volatility in the market. On the
other hand, some argue that greater liquidity trading by institutions reduces spread.
There is no direct empirical evidence and little theoretical knowledge to suggest a
convincing relation between institutional trading and spread. In this paper, we present
some evidence on the nature and effect of institutional trading on spreads. We argue that
institutional trading is not completely information driven, part of it is liquidity trading in
nature. We find evidence that information induced institutional trading increases the
adverse selection component. However, large volume (liquidity) trading reduces the
order processing costs. We find the net effect of institutional trading on spread is
consistently negative. Moreover, institutional buys have differential information from
sells. Institutional trades per se reduce spreads, but only sells increase the adverse
selection component. Both effective and relative spreads impound the differential nature
of institutional buys and sells.
Keywords: Institutional Trading, Spread decomposition, information, liquidity.
JEL Classification: G19.
Institutional trading, Trading Volume, and Spread
Since Demsetz (1968) bid-ask spread is recognized as the price of liquidity
provided by the dealers in an equity market. A number of studies have investigated what
determines spread (Branch and Freed , McInish and Wood , Klock and
McCormick , Heflin and Shaw ). Some of the significant determinants of
spread found in the literature are order size, number of trades, competition in the dealers’
market, ownership structure, and the native characteristics of a stock e.g., price, and
volatility. Trading rules and mechanics of trading that proxy for information flow are
also found to affect the spread.
There is very little empirical evidence on institutional trading and spread and their
interrelationship. Keim and Madhavan (1997) find execution costs for institutional trades
are different between listed and NASDAQ stocks. Conrad, Johnson and Wahal (2001)
report an asymmetric relation between institutional buys and sells and soft-dollar
execution. However, there is some evidence of the effect of institutional trading on
securities prices. Empirical studies using order size or trading volume as a proxy for
institutional trading1 suggest an increased price effect associated with institutional
trading. Using proprietary data on institutional trading, Chan and Lakonishok (1993) find
the average price effect to be small for institutional trades, but the price effect for buys
and sells to be asymmetric; and Sias and Starks (1997) find that institutional trading
contributes to serial correlation of returns.
1 Lakonishok and Maberly (1990) use block trades as a proxy for institutional trades, and odd lot trades
plus non-margin account trades as a proxy for individual trades.
The role of institutional trading in the determination of spread is interesting since
it is often argued (casually) that increased institutional participation in the U.S. equity
market during the past decade has led to an increase in the volatility, and has widened the
bid-ask spread in the equity market. On the other hand, some argue that institutional
trades provide liquidity, and hence decrease spread. Bertisimas and Lo (1998) show how
optimal trading strategies may be devised by execution cost minimizing investment
mangers. In any case, there is very little empirical evidence or theoretical knowledge to
conclude how institutional trading affects spread. Further since spread is considered to
be a sum of two different components, adverse selection, and order processing2 it is
unclear how institutional trading affects the individual components of the spread.
In this paper, we investigate if institutional trading has any information content
beyond what has been documented as a size or volume effect. In a multivariate, panel
regression framework, we determine if there exists a relation between bid-ask spread and
institutional trading after adjusting for size and price effects. Most studies on the
determinants of spread focus on the supply side of dealership market i.e., competition in
the dealer market, and use a cross sectional regression approach. Our approach is
different from previous studies in two important ways. First, we focus on the demand
side (investors’ characteristics) to determine the relation between spread and institutional
trading3; and second, we use a panel data approach accounting specifically for both serial
and contemporaneous correlation in the error terms4. We report regression results using
2 Arguably there is also inventory effect in spread. In this paper, we use a decomposition technique that
ignores inventory effect.
3 Recently, some other papers have studied the demand side, particularly the ownership structure. For
example, Heflin and Shaw (2000) document a relation between block holding and spread, and Chung and
Charoenwong (1998) look at insider holding as a determinant of spread.
4 Dey (2000) introduces a similar but reduced form panel data regression model for effective spread.
both effective ($) and relative spread as the dependent variable in a set of regression
Further we decompose the spread into order processing and adverse selection
components and investigate how those components vary with changes in trading volume,
net order flow (buy vs. sell), and institutional trading. We assume contemporaneous
correlation between the disturbances and use an SUR (Seemingly Unrelated Regression)
analysis to find the significant determinants of the adverse selection and the order
processing components of the spread for our sample firms. We use a unique data set
(TORQ) that identifies institutional trading. Prior studies proxy institutional participation
by using measures based on trade size that are subject to measurement error.
Our results show that institutional trading proportion is inversely related to both
effective and relative spreads. We also find that the negative slope (suggestive of the
inverse relation) is not constant and flattens out at higher concentrations of institutional
trading. We find that this negative slope is provided by both institutional buys and sells
alike. Results from a SUR analysis show that the adverse selection costs tend to increase
and the order processing costs tend to decrease with increases in institutional trades.
The rest of this paper is organized as follows. In section II, we present the
motivation for this study. In section III, we describe an empirical model for spread using
panel regressions, provide data description, and explain the results. Further, we describe
the decomposition of the spread into order processing and adverse selection costs
components and report results from a SUR analysis of the determinants of those
components. Section IV concludes the paper.
II.A. Relation between Spread and Institutional trading
Schwartz (1988) identifies four classes of variables, namely, activity, risk,
information, and competition as determinants of spread. Existing literature find trade
size, number of trades, ownership structure, and extent of market power in the dealership
market to be the key determinants of bid-ask spread (McInish and Wood , Laux
[1993, 1995], Klock and McCormick , Heflin and Shaw ). Dealer market
competition represents the supply side of the market for liquidity services5. On the other
hand, trade size, ownership structure, and frequency of trading measure the activity in
securities markets and represent the demand side of the market for liquidity services.
Prior research suggests an inverse relationship between spread and trading activity
measured by order size, and number of trades (McInish and Wood ). Institutions
trade large sizes, and also trade frequently.6 Thus institutional trading will induce low
spread. However, trading activity also contributes to both information and risk associated
with a security7. Hasbrouck (1991) provides evidence that large trades contain more
information than small trades and cause spreads to widen. Lin, Sanger and Booth –
hereafter LSB (1995) find evidence of an increasing (although not continuously), non-
linear relation between spread and trade size. Jones, Kaul, and Lipson (1994) report that
5 In a securities market, dealers provide liquidity (both buying and selling of securities) services, while
investors demand those services.
6 Lee and Radhakrishna (2000) confirm earlier evidence that institutions tend to trade in larger volumes.
They present a technique of classifying trades into institutional and non-institutional based on size in a way
that reduces the error in classification to statistically manageable levels.
7 See, for example, the summary of the empirical evidence on the price effect of block trades in Dey (2000).
the number of trades captures the essence of volatility in financial markets even in the
presence of volume and trade size.8
Seppi (1990) argues that institutions use trade size strategically; they trade large
orders when they can signal to the market that their trades are not information motivated
and hence large institutional trades may not have information content. Dey and Kazemi
(2000) distinguish between large, pure information trades and large institutional trades
and argue that institutional trades are driven by both “pure information” and liquidity
needs. Dey and Kazemi (2000) predict the “pure information” component of equilibrium
spread to be an increasing, while the liquidity component of the spread to be a decreasing
function of institutional trading.
Chan and Lakonishok (1993), and Keim and Madhavan (1994) find that the price
effect and cost implications of institutional buys and sells are not symmetric. Koski and
Michaely (2000) find that buys and sells provide different information for different trade
sizes. Saar (2000) provides a theoretical framework based on a dynamic portfolio
rebalancing process of institutions to explain the documented asymmetry in the price
effect of institutional buys and sells.
In this paper, in a multivariate regression framework, we determine whether
institutional trading has information content beyond that provided by trade size denoted
by trading volume, and number of trades. Specifically, we hypothesize that the variation
in bid-ask spread can be explained by trading volume, number of trades, price, and
institutional trading. Further, we hypothesize that institutional trading per se and not the
8 Recently, Chan and Fong (1999) find that after adjusting for order imbalance, trade size is important for
the volume-volatility relation.
direction of institutional trading - buy or sell affects the bid-ask spread. Stated in
alternate form, we hypothesize:
H1: The bid-ask spread should vary significantly with institutional trading
after controlling for number of trades, price, and trading volume. Further,
institutional trades per se affect bid ask spread, and thus institutional buys
and sells do not have any differential effect on the bid-ask spread after
controlling for number of trades, price, and trading volume.
II.B. Components of the spread
We extend our analysis of the relation between institutional trading and spread by
decomposing the spread into its order processing and adverse selection components and
investigating the effect of institutional trading on the individual components. We use a
technique from LSB (1995) to decompose the spread into order processing and adverse
selection components and hypothesize a relation between the individual components and
Sias (1996) reports positive correlation between institutional activity and market
volatility; however, Cohen et al (1987) suggest that institutions trade frequently because
of their low order processing costs. Bertisimas and Lo (1998) derive optimal trading
strategies of institutions based on minimum execution cost. LSB (1995) find the order
processing cost to be decreasing and adverse selection cost to be increasing in trade size.
We conjecture that if institutional trading is a mix of information and liquidity trading
then the information effect should increase the adverse selection component and the
liquidity effect should decrease the order processing costs. We also recognize that while
gross volume is important for the determination of order processing cost, trade direction
or net volume (buy volume - sell volume), is important in the determination of adverse
selection costs. We therefore include log (buy/sell) as a variable in the regression model
for the adverse selection component.
We determine through a set of simultaneous equations how institutional trading
affects the order processing and the adverse selection components of the spread after
controlling for number of trades, volume, and trade direction. The simultaneous
equations approach uses the cross correlation between the two regression equations to
improve the estimates. Further we determine how the asymmetric information content
and the liquidity motive in institutional buys and institutional sells affects the adverse
selection (information) component of the spread. Stated in alternate form, our hypotheses
H2a. The adverse-selection component should increase with institutional
trading, and the order-processing component should decrease with
H2b. Institutional buys should have a differential effect on the adverse
selection component of the spread from institutional sells.
III. Regression Models, Data Description, and Results
III.A.1. Panel Data Regression Model
We use a multivariate, panel (time series – cross sectional) regression framework
to investigate the effect of institutional trading on the bid-ask spread. We use a unique
data set (TORQ) that allows us to identify the order origination for a trade as institutional
or otherwise. Most studies on the determinants of spread use pooled OLS estimates of
the parameters of a regression model. OLS estimates ignore the covariance structure of
the error term both across firms and over time.
We assume disturbances are both serially and contemporaneously correlated.
Specifically, we assume an AR(1) process with contemporaneous correlation for the
disturbance term. In our model for the spread, the serial correlation may be due to lagged
spread or lagged values of the independent variables or their interactions. Kim and
Ogden (1996) find higher order serial correlation for the spread, and Peles (1992) report
contemporaneous correlation among equity trading of institutional investors. Parks
(1967) provide consistent and efficient estimates of the parameters when disturbances
follow a first order auto regressive process - AR(1) with contemporaneous correlation.
We run the following regression model for our panel data:
n = 1…N; number of firms in sample, t = 1…T; number of trading days
Spreadnt = Effective or Relative spread for the nth firm on day t
NtradePctnt = Number of trades on day t for firm n expressed as a proportion of
average number of daily trades over the entire sample period
Dly_Avgnt = Volume of trade on day t for firm n expressed as a proportion of
average daily trading volume over the entire sample period
Pricent = Closing price on day t for firm n
Instpropnt = Institutional trading (number of trades) as a proportion of total
trading for the nth firm on day t.
Further, the error structure is assumed as follows:
e r em
ntn n tnt
and ,(due to contemporaneous correlation),
(disturbance term is AR(1)),
, 0 b
0 where (cross correlation is zero).
To estimate the parameters of our regression model, we use data from the TORQ
data set. The TORQ files released by NYSE were prepared under the supervision of
Professor Joel Hasbrouck during his tenure as a Visiting Economist to the NYSE. This
dataset contains trades, quotes, order processing, and audit trail data for a sample of 144
NYSE stocks for the three months (63 trading days) from November 1990 through
January 1991.9 These firms represent a size stratified random sample of firms in the
NYSE and thus cover the broad spectrum of NYSE firms.
As noted by Lee and Radhakrishna (2000), the marginal contribution of TORQ
data over ISSM or TAQ data is in providing identification for traders’ classes, as
institutions, individuals, and dealers. Most studies using other trades/quotes databases
use size as a proxy for institutional trades.
We impose a restriction that is common among studies that study effective
spreads or the components of the spread to ensure adequacy of data in estimation. We
select all firms in the data set that have on average 20 trades per day or more during the
sample period. Further, in classifying the trades, whenever there are executions of
multiple orders on the active side of a trade, we take the trader class of the largest order
9For more details on the TORQ database and on trading procedures at the NYSE, see Hasbrouck (1992).
on the active side as the initiator of that trade. The active side of the trade is determined
using the Lee-Ready (1991) algorithm.10 This reduces our sample to 65 firms.
Of the 65 firms that survived our initial cutoff, 14 firms have one or more days of
missing observations or days with trading halts. We leave out the firms with trading halt
days from our study since the effect of trading halts on spreads and the price discovery
process is unique and beyond the realm of this paper. We chose to omit the firms with
missing observations, since there are questions about the reliability of estimates using an
unbalanced panel11. Thus we have a balanced panel of 51 firms with 63 days data that we
use in our panel regression.
The panel data set includes daily data for the firms in the data set. For each firm,
we calculate the mean daily effective bid-ask spread, and the proportion of buy and sell
orders initiated by institutions for each day. We compute the effective spread for each
trade defined as twice the absolute value of the difference between trade price and the
prevailing mid-quote. The mean effective spread is the average of effective spreads
across all trades in a day. To determine the mean proportion of trades by institutions in
that firm, we calculate, for each day, the proportion of trades by institutions on the buy
and sell side.
Besides proportion of institutional trading, there are three other independent
variables in the panel regression. The “number of trades” variable is computed as the
number of trades each day divided by the average number of trades for the firm over the
10 Lee and Radhakrishna (2000) show the effectiveness of Lee-Ready algorithm for classifying trades into
buys and sells.
11 We explored another alternative i.e., to consider all the firms (65) but for fewer days (61). However,
there was a problem in deciding which 2 days to delete for firms that have full 63 days data. Since any
choice on this was likely to be arbitrary, we chose to use a balanced panel of 51 firms over 63 days.
sample period. Thus this is a measure of abnormal trading in each day. Trading above
(below) mean would give this variable a value higher (lower) than one. The “trading
volume” variable is computed by dividing the daily share volume for the firm by the
average daily share volume over the sample period. Therefore, this variable also has
values above (below) one when trading volume is higher (lower) than the average daily
share volume. The price variable is the closing price of the stock.12
In Table 1, we present means of the computed statistics of the variables used in
the panel regression. We first compute the relevant statistics for the sample firms over
the sample period and then compute the means of those statistics. Thus we report the
means of the cross sections of firm means, medians, and standard deviations. The inter-
firm mean (median) spread is .126 (.12) that is about an eighth. The inter-firm mean
standard deviation is quite low at .02. The largest spread in our sample is .514,
approximately one-half, and the lowest .019 or approximately one-sixty-fourth. The
inter-firm mean (and median) institutional trading in our sample is around 30%.
Although institutions generally trade on a regular basis, in some trading days there is no
institutional trading. Of the 3,213 (51×63) firm trading days covered in our sample there
were 10 firm-days when there was no institutional trading, and one firm-day when all the
trading was initiated by institutions. There is also sufficient variability in the number of
trades and volume. The low trading (volume) in our sample was 9% (3%) of average
daily trades (volume). The high trading (volume) was 392% (1235%) of average daily
volume. In Table 2, we present Pearson correlations for variables in the regression.
12 Using average price over a day instead of closing price does not change the results.
Panel A presents pooled correlations computed from 3,213 observations. In Panel B, we
present the means of correlations computed in time series for each firm.
III.A.3 Regression Results of Panel Model
Table 3 reports the regression results for the panel data regressions with effective
spread (in dollars) and relative spread as dependent variables. For our first model (1-ES),
the independent variables are number of trades, trading volume, price, and the proportion
of institutional trades. All four variables are significant in determining effective spread.
The significant coefficients show that effective spread increases as trading (number of
trade) and price increase and decreases as trading volume and institutional trading
increase. The coefficient for institutional proportion is negative (-.0111) and significant at
less than 1% level. Thus an increase in institutional proportion reduces the spread. The
R2 for this model is 22%. However, for a similar model with relative spread (1-RS) as
the dependent variable, number of trades fails to be a significant determinant of spread.
All other variables, namely average trading volume (-.0137), price (-.028), and
institutional proportion (-.024) remain significant at less than 1% level, and the R2 for the
model is 90%. The change in the effect of price on relative spread (decreasing in price)
from that on effective spread (increasing in price) is expected since relative spread is
computed as effective spread over mid quote. This change in the effect of price on
effective and relative spreads is consistent across all the four models.
For our second model, we introduce two dummy variables for high and medium
institutional proportions. The high (medium) institutional proportion variable has a value
of 1 when the level of institutional trading is in the top (middle) 33% percentile, and zero
otherwise. The coefficients of the dummy variables are both positive and significant in
the regression. The high dummy has a larger coefficient (.0016) than the medium
dummy variable (.0007). Taken in conjunction with the significant negative (-0.0153)
coefficient of the institutional proportion variable, this suggests that the negative slope of
the institutional proportion variable flattens out at higher levels of institutional trading.
This conjecture is confirmed with model 3.
In model 3, the institutional proportion variable is replaced by three variables –
high, medium and low proportion. The high (medium, low) proportion variable has the
same value as institutional proportion if institutional trading proportion is in the top
(middle, bottom) 33 percentile, and zero otherwise. The coefficients of all three variables
are negative and significant in the regression, but while the coefficient for low proportion
is -.0144, that for the high proportion is significantly (14%) less at -.0122. We interpret
these results as follows. On average, there may be a mean positive effect on institutional
trading embodied in the positive intercept. However, when there is an increase in
institutional trading within a level, the spread declines, but the rate of decrease is lower at
higher levels of institutional proportion. Between low and medium levels, the rate of
decrease is similar indicating these are perhaps two discrete levels of institutional
Finally, for our fourth model, we break up the institutional proportion into
institutional buy and institutional sell to test for a difference in their effect on the spread.
Results from the fourth model show that spread reduces as institutional trading increases,
be it buy or sell. The coefficients for buy and sell are significantly negative for both
effective spread and relative spread.