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Published in: Review of Quantitative Finance and Accounting 3/2016

01-10-2016 | Original Research

When noise trading fades, volatility rises

Author: Jinliang Li

Published in: Review of Quantitative Finance and Accounting | Issue 3/2016

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Abstract

We hypothesize and test an inverse relation between liquidity and price volatility derived from microstructure theory. Two important facets of liquidity trading are examined: volume and noisiness. As represented by the expected turnover rate (volume) and realized average commission cost per share (noisiness) of NYSE equity trading, both facets are found negatively associated with the ex post and ex ante return volatilities of the NYSE stock portfolios and the NYSE composite index futures. Furthermore, the inverse association between noisiness and volatility is amplified in times of market crisis. The negative noisiness–volatility relation is also supported by our analysis on the effects of trade size on price volatility. The overall results demonstrate that volatility increases as noise trading declines.

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Appendix
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Footnotes
1
Noise in the sense of a large number of small events is often a causal factor much more powerful than a small number of large events can be (Black 1986). The terms “noise trading” and “liquidity trading” are used interchangeably in the market microstructure literature. We use “noise trading” to emphasize the nosiness and “liquidity trading” to emphasize the volume of liquidity.
 
2
This view is consistent with the work of Huang and Wang (2009), which postulates that endogenous illiquidity can lead to market crashes in the absence of any aggregate shocks.
 
3
Shleifer and Summers (1990) discuss arbitrage limits and conclude that “news alone does not move stock prices; uninformed changes in demand move them too.” Irrational noise traders with erroneous stochastic beliefs may affect prices and earn higher expected returns (De Long et al. 1990). In the presence of heterogeneous beliefs, Daigler and Wiley (1999) and Shalen (1993) have found that uninformed trading may cause price volatility.
 
4
Brennan and Subrahmanyam (1996) combine diverse empirical techniques from asset pricing and market microstructure research to study this relation. They find the microstructure parameters of market depth and illiquidity to be related to stock return premia. They relate illiquidity to adverse selection costs, which are costs associated with the discovery of private information that are paid by liquidity traders.
 
5
Johnson (2008) provides a model in which the equilibrium price-response measure of liquidity reflects the average risk-bearing capacity of the economy and volume reflects the changing contribution of individuals to that average. In that model, volume and liquidity are unrelated but jointly determined.
 
6
By AIC and BIC, an AR(1) process with a time drift is adequate to fit the logarithm quarterly turnover rate.
 
7
French et al. (1987) use one lagged cross-variance in Eq. (2) and make no adjustment for the mean return. We also estimate the quarterly volatility following their method. The estimates following French et al. (1987) are all positive and very close to the estimates in Eq. (2). The empirical analysis using the French et al. estimates of quarterly volatility is consistent with the analysis using the estimates in Eq. (2). For brevity, we do not report the analysis using the French et al. estimates.
 
8
According to Eq. (3), the association between \(\widetilde{{\Delta {\text{ACR}}_{\text{t}} }}\) and quarterly volatility (variance) can be written as \(\sigma_{\text{t}}^{2} = \exp \left( {2\varphi_{1} \cdot \widetilde{{\Delta {\text{ACR}}_{\text{t}} }} } \right) \cdot \exp \left[ {2\left( {\mu + \sum\nolimits_{{{\text{m}} = 2}}^{\text{M}} {\varphi_{\text{m}} \cdot {\text{f}}_{{{\text{m}},{\text{t}}}} + \frac{{1 - \theta {\text{L}}}}{{1 - \rho {\text{L}}}}\varepsilon_{\text{t}} } } \right)} \right]\), where \({\text{f}}_{{{\text{m}},{\text{t}}}}\) refers to the explanatory variables in addition to \(\widetilde{{\Delta {\text{ACR}}_{\text{t}} }}\). The marginal proportion of the volatility that is explained by \(\widetilde{{\Delta {\text{ACR}}_{\text{t}} }}\) is \(\exp \left( {2\varphi_{1} \cdot \widetilde{{\Delta {\text{ACR}}_{\text{t}} }} } \right) - 1\).
 
9
We also run the regressions with restrictions of the GARCH parameters and yield essentially identical results.
 
10
In the following reported GARCH regressions, we include the MA term. We also run the GARCH regressions without this MA term, and yield qualitatively similar results.
 
11
It is well known that the residual series in the GARCH process is strictly stationary with finite second moment provided α + β < 1 (see Tsay 2005: page 14). Engle and Rangel (2008: page 1191) explicitly point out that when α+β<1 , the conditional variance reverts to its mean value at a geometric rate of α + β.
 
12
\(\widetilde{{\Delta {\text{ACR}}_{\text{t}} }}\) is related to the variance of daily stock price returns in the following way: \(\sigma_{i,t}^{2} = \exp \left( {\varphi_{1} \cdot \widetilde{{\Delta {\text{ACR}}_{\text{t}} }} } \right) \cdot \exp \left( {m + \sum\nolimits_{n = 2}^{N} {\varphi_{n} \cdot f_{n,t} } } \right) \cdot h_{i,t}\), where \(\varphi_{n}\) (n = 1 to N) denotes the coefficient of \(\widetilde{{\Delta {\text{ACR}}_{\text{t}} }}\) or other corresponding liquidity variables in the regressions. The marginal (proportional) effect of \(\widetilde{{\Delta {\text{ACR}}_{\text{t}} }}\) on return variance is thus \({ \exp }(\varphi_{1} \cdot \widetilde{{\Delta {\text{ACR}}_{\text{t}} }}) - 1\).
 
13
This paper focuses on “micro” liquidity (trading volume and transaction costs), while the government often refers to “macro” liquidity (money supply in general or fund flows between asset sectors).
 
14
Jones (2002) reports a similar estimate.
 
15
Report that brokers charge a per-share commission to institutional traders as a convenient way of charging a predetermined fixed fee for broker services.
 
16
We use the 30 DJIA component companies as of the beginning of 2004. They are 3 M, Alcoa, Altria Group, American Express, AT&T, Boeing, Caterpillar, Citigroup, Du Pont, Eastman Kodak, EXXON Mobil, General Electric, General Motors, Hewlett–Packard, Home Depot, Honeywell International, Intel, IBM, International Paper, Johnson & Johnson, J. P. Morgan Chase, Coca-Cola, McDonald’s, Merck, Microsoft Corporation, Proctor & Gamble, SBC Communications, United Technologies, Wal-Mart Stores, and Walt Disney. AT&T, International Paper, and Eastman Kodak were subsequently replaced by AIG, Pfizer, and Verizon on April 8, 2004. However, AIG was later removed from the index on September 22, 2008.
 
17
NYSE TAQ databank traces back to the beginning of 1993.
 
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Metadata
Title
When noise trading fades, volatility rises
Author
Jinliang Li
Publication date
01-10-2016
Publisher
Springer US
Published in
Review of Quantitative Finance and Accounting / Issue 3/2016
Print ISSN: 0924-865X
Electronic ISSN: 1573-7179
DOI
https://doi.org/10.1007/s11156-015-0508-2

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