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1 Chinese institutional investors’ sentiment Gerhard Kling a,*, Lei Gao b a University of Southampton b School of Business, Shantou University, Shantou 515063, PR China THIS IS NOT THE FINAL (POST-REVIEW) VERSION YOU FIND THE FINAL VERSION HERE: Kling, G. and L. Gao (2008) Chinese institutional investors' sentiment, Journal of International Financial Markets, Institutions & Money 18(4), 374-387. We use daily survey data on Chinese institutional investors‟ forecasts to measure investors' sentiment. Our empirical model uncovers that share prices and investor sentiment do not have a long-run relation; however, in the short-run, the mood of investors follows a positive feedback process. Hence, institutional investors are optimistic when previous market returns were positive. Contrarily, negative returns trigger a decline in sentiment, which reacts more sensitively to negative than positive returns. Investor sentiment does not predict future market movements – but a drop in confidence increases market volatility and destabilizes exchanges. EGARCH models reveal asymmetric responses in the volatility of investor sentiment; however, Granger causality tests reject volatility-spillovers between returns and sentiment. 4th April 2007 JEL classification: G14; C22 Keywords: Shanghai stock exchange; Institutional investor; Investors‟ sentiment * Corresponding author E-mail addresses: drgaolei@126.com (L. Gao), g.kling@soton.ac.uk (G. Kling). 2 1. Introduction Research on investors‟ sentiment has focused on small shareholders for two main reasons: (1) data on the sentiment of large investors are not available; (2) small shareholders are supposed to act like noise traders; thus, the impact of noise trading can be analyzed.1 In contrast, we measure the sentiment of institutional investors in China by collecting daily survey data on investors‟ forecasts. Hence, we know how many institutional investors expect a positive or negative change in the Shanghai Stock Exchange (SSE) index on the next day. The interrelation between institutional investors‟ sentiment and stock markets might differ from former findings, as institutional investors are more professional in analyzing public information, conduct large-scale transactions, and possess informational advantages. In China, several authorized closed-end investment funds exist, and open-end funds were recently introduced. Yet other forms of institutional investment like pension funds or managed portfolios of insurance companies are negligible. Based on World Bank data for 2001, Chinese institutional investors‟ assets reached 19% of GDP; this figure was lower than the corresponding figures for Hungary (26%), the Czech Republic (32%), and further developed Asian markets, e.g. South Korea (82%). As Chinese institutional traders are less important than their US and European counterparts, one can regard our analysis as a lower estimate of the importance of institutional investors‟ sentiment in other markets. Three questions are at the core of our interest: (1) what determines the forecasts of institutional investors; (2) can the sentiment index predict future stock market movements; (3) does a drop in sentiment or a higher volatility of investors' sentiment destabilize stock markets by triggering higher market volatility? Besides these core issues, we analyze asymmetric responses of investor sentiment based on EGARCH and TARCH models. In addition, 1 Kelly (1997) showed that the likelihood of being a noise trader declines with household income; thus, the analysis of noise trading is limited to small investors. We define noise trading in the sense of DeLong et al. (1990). 3 bivariate GARCH models allow us to explore the time-varying correlation between sentiment and stock returns. The literature on investors‟ sentiment covers our research questions; however, only the sentiment of small investors has been studied so far. Our first question, the search for factors that affect investors‟ sentiment, is still under discussion – albeit the formal model of investor sentiment developed by Barberis, Shleifer, and Vishny (1998) has provided a baseline for economists. The literature focused mainly on our second question: Neal and Wheatley (1998), for instance, quantified the predictability of stock returns using different measures of investors‟ sentiment. Lee, Jiang, and Indro (2002) discussed our third question for the US in analyzing the influence of investors‟ sentiment on volatility and excess returns. The remainder of our paper is organized as follows: section two reviews the recent literature on theoretical considerations concerning sentiment and market activity, institutional trading, sentiment of small investors, survey methods, and indirect measures of sentiment. Section three describes our data sources, followed by our empirical analysis that starts with a univariate-time series approach followed by unit-root tests and multivariate methods, namely VAR and bivariate GARCH models. Based on our empirical findings, we discuss consequences for mature financial markets and peculiarities of the Chinese stock exchange. 2. Literature review Classical finance theory neglects the role of investor sentiment (see Gomes, Kogan, and Zhang, 2003), as investors are supposed to be rational. Even if some investors are not rational, arbitrageurs can exploit their irrational behavior, thus causing prices to reflect future discounted cash flows. By introducing demand shifts driven by irrational speculation and binding arbitrage constraints, sentiment can influence stock returns. Hence, a mispricing triggered by uninformed demand shocks could occur. This is in line with the idea that sentiment can be regarded as the propensity to speculate and thus reflects the optimism or 4 pessimism of investors. Especially if arbitrage is risky because of subjective valuations, 2 high volatility, short selling constraints, 3 and thin trading, the impact of sentiment on returns should be more pronounced. In particular, the situation in China is characterized by the lack of an earnings history, apparently unlimited growth potential, and unsophisticated investors; hence, the role of investor sentiment should be tremendous. As institutional investors account for more than 50% of trading in the US (see Cai and Zheng, 2004), their influence on stock markets is of considerable interest. Previous studies have mainly focused on the impact of institutional trading on stock returns, and assessed the consequences of large-scale transactions, positive-feedback trading, and the herding behavior of institutional investors (see Lakonishok, Shleifer, and Vishny, 1992). Nevertheless, Bohl and Brzeszczynski (2006) studied the influence of institutional ownership on volatilities. They found that institutional investors stabilize exchanges, as volatility drops after institutional investors become principal shareholders. These studies, however, have in common that they analyze the effect of institutional trading and ownership, whereas our paper tries to measure the relevance of institutional investors‟ sentiment for stock returns and volatilities. This requires data on institutional investors‟ sentiment, which are usually not available as surveys or alternative measurement methods focus on small shareholders‟ sentiment. Results on whether the sentiment of small investors is a good indicator of future market movements are somehow ambiguous. Brown and Cliff (2001) provided only weak evidence of short-run predictability, but found long-run predictability of returns for a period of about two to three years. Kenneth and Statman (2000) detected a significantly negative correlation between US investors‟ mood and returns of the S&P500. Apart from the 2 A broad range of valuations increases noise trader risk (see DeLong, Shleifer, Summers, and Waldmann, 1990; Shleifer and Vishny, 1997). 3 Arbitrage is limited by short selling constraints (see e.g. Jones and Lamont, 2002). In China, the short selling of stock is hardly possible and derivate instruments are not developed.. 5 discussion about whether investor sentiment can predict future market movements, several studies identified an interrelation with market volatility. By using GARCH models, Lee, Jiang, and Indro (2002) analyzed the impact of investors‟ sentiment on market volatility. They uncovered that a decline in investors‟ sentiment triggers higher volatility. Several direct and indirect measures of sentiment have been used in the recent literature. For US studies, the sentiment index provided by Investors‟ Intelligence of New Rochelle based on a weekly survey of 135 independent advisory services, is commonly applied. 4 Survey methods have provoked the skepticism of many economists. Yet Blinder (1991) argued that a well-designed survey can provide stylized facts and valuable clues that go beyond standard models and are not available to econometricians. The so-called indirect approach serves as an alternative to survey data. For US studies, the fluctuations of closed- end fund shares are used as a proxy of sentiment because predominantly private investors hold these shares.5 Besides observing closed-end funds, Baker and Stein (2004) argued that market liquidity could be regarded as a proxy of investors‟ sentiment. They used bid-ask spreads and turnovers as a measure of liquidity. This measure refers to the whole market; thereby, it is impossible to distinguish between different types of investors. Consequently, we favor a survey approach in order to isolate the opinion of institutional investors and their market impact from all other market participants. 3. Data Since April 20, 2001, the Chinese Central Television Station has been surveying 75 leading institutional investors. On a daily basis, they are asked about their forecast regarding the Shanghai Stock Exchange Index‟s movements on the following day. The survey is conducted 4 Many studies like Siegel (1992) used this index and showed that it can predict market movements. 5 Lee, Shleifer and Thaler (1991) proposed this method, and Neal and Wheatley (1998) showed that fluctuations of closed-end funds can predict the returns of small caps. 6 at 4pm each day, and the results are published online. 6 As the market closes before 4pm, institutional investors could use today‟s market development as information for their forecast. The surveyed institutional investors have three choices: bullish, bearish, and tie. The web page contains the numbers of investors opting for each type of forecast – but individual opinions remain confidential. Based on these figures, we calculate for each day the percentage of optimistic institutional investors and set up the daily index of optimism. This index is constructed in a similar way as the index provided by Investors‟ Intelligence of New Rochelle, which is used for US studies. The sentiment index reaches 0 if everyone expects a decline in the stock market, and 100 if everyone is optimistic. Henceforth, the sentiment index is bounded between 0 and 100; however, the variable is normally distributed, as extreme cases of 0% or 100% optimistic responses are rare.7 Three major aspects make the Chinese daily survey unique: first, the survey is based on the opinion of institutional investors, which allows the measurement of their sentiment. Second, it provides daily information on investors‟ mood and thus has the highest possible frequency of information. Sentiment indices for the US are usually on a weekly basis. Third, the forecasting horizon is only one day, as investors are asked about their forecast regarding tomorrow‟s market movement. Generally, there is no hint that the survey we use contains any incentives to give false information, as individual forecasts remain confidential. 4. Empirical models 4.1. Univariate analysis of investor sentiment Daily institutional investors' sentiment fluctuates remarkably over time (see Fig.1). To model the variance pattern, we use an ARCH (p) model based on the seminal paper of Engle (1982). 6 See http://www.eestart.com/zxzx/20040220/558105.html. 7 Shapiro-Wilk and Shapiro-Francia W tests indicate, with p-values of 0.084 and 0.122, that the null hypothesis of the sentiment index being normally distributed cannot be rejected at the 95% level of confidence. 7 The log of the sentiment index is labeled yt, and its conditional variance is t2. Hereafter, we always refer to the logarithmically transformed sentiment index except when stated otherwise. yt 0 t (1) E t t 0 Var t t t2 0 j t2 j p j 1 To determine the appropriate lag structure of the ARCH (p), we conduct LM tests by regressing squared residuals of equation (1) on their lagged values. 8 Justified by these outcomes, we specify an ARCH (2) process. Besides dealing with a nonlinear structure in conditional variances, one should capture the autoregressive behavior of the sentiment index. Accordingly, the next step is to specify an ARMA (p, q) model and insert this structure into the mean equation (1).9 By looking at the autocorrelation function and the partial autocorrelation function, we can justify an ARMA (3, 4) specification. A reduction of the lags does not lower information criteria like the Akaike criterion; thus, we incorporate an ARMA (3, 4) term into our mean equation. 3 4 (2) yt 0 j yt j j t j t j 1 j 1 2 t2 0 j t2 j j 1 Fig.1 depicts the actual and predicted values of the sentiment index. To ensure that our model captures all nonlinearities in the time series, we confirm through white noise tests that the 8 Multiplying the number of observations and the R-squares achieved for different lag structures yields a test statistic that is Chi-squared distributed with p degrees of freedom. It is not possible to reject the null hypothesis that ARCH terms are jointly insignificant for an ARCH (3) specification and higher orders – but with a p-value of 0.001 for the Chi-squared test, an ARCH (2) specification has significant coefficients. 9 To reliably estimate an ARMA specification for the sentiment index, the time series has to be stationary. By looking at Fig.1, one obtains the impression that this requirement is fulfilled. Also, ADF tests indicate that the time series does not possess a unit root. A subsequent section discusses this issue more thoroughly. 8 residuals exhibit erratic behavior.10 The Barlett‟s periodogram-based test procedure fails to reject the null that the series is white noise, and the Portmanteau Box-Pierce test points in the same direction.11 How can one interpret these findings? The negative autocorrelation suggests that a positive forecast is likely to be followed by a negative one. Yet the importance of former assessments dies out quickly. The following section focuses on the interrelation between sentiment, stock returns, and volatility. 4.2. The interrelation between sentiment and stock returns: a VAR analysis Before using more elaborate techniques, a simple cross-correlogram can help to detect whether returns or forecasts move first. Fig.2 plots the cross-correlation for different lags and leads. A positive correlation of about 0.4106 between lagged market returns and forecasts reveals an interesting interdependency. Previous market returns influence the current mood of institutional investors positively. This is a highly interesting finding, as current market returns do not affect forecasts despite the fact that the survey is conducted after the market is closed. Institutional investors could use current returns to improve their forecasts, yet they rely on past information. To study this time structure and correlation further, our strategy is as follows. First, by applying unit-root tests, we test whether both time series are stationary. Second, we model the short-term interrelation by a VAR approach and select the lag structure based on information criteria. Thereafter, Granger causality tests show whether forecasts affect market returns or vice versa. 10 Before performing white noise tests, one has to fill two gaps in the time series of the residuals. Because the problem of missing values is rather limited, we propose a very simple solution for this problem. According to this, the two missing values in the time series are replaced by the previous values of the residuals. 11 The Barlett‟s test statistic reaches 0.950 (p-value: 0.327), and the Portmanteau Q statistic is equal to 39.622 (p- value: 0.487). 9 We test for stationarity by calculating ADF based tests. 12 Due to the „traditionally‟ weak explanatory power of ADF tests, we also report the results of KPSS tests that use a different null hypothesis.13 If both test procedures point in the same direction, the result can be regarded as informative. Both test procedures point in the same direction and thus confirm that both time series are stationary.14 As investor sentiment is I(0) and stock prices are I(1), both time series cannot be cointegrated and a long-run relationship does not exist. However, we analyze the short-run dynamics by a VAR model and determine the number of lags as suggested by information criteria, namely Schwarz Bayesian (SBIC), Akaike (AIC), and Hannan-Quin (HQIC). As expected,15 the SBIC favors the inclusion of two lags, whereas AIC and HQIC request one additional lag. Hence, we estimate a VAR with two and three lags, respectively, and carry out Granger causality tests. Table 1 presents the outcomes. The interpretation is twofold. Only the market return rt-1 of the previous day affects the current forecasts yt. Moreover, Granger causality tests reveal that investor sentiment does not influence returns and hence is not a valuable information for the market. It is noteworthy that a positive market return triggers a positive forecast of the institutional investors. Accordingly, institutional investors‟ sentiment can be interpreted as a positive feedback process with regard to previous market returns. Variance decompositions highlight that an impulse in returns determines about 20% of the subsequent innovations in investor sentiment. 12 Elliott, Rothenberg, and Stock (1996) modified ADF tests by a GLS approach that accounts for serial correlation. 13 Kwiatkowski et al. (1992) constructed this test procedure. 14 The ADF test statistic reaches -4.511 for the sentiment index and -17.451 for market returns. Hence, the null hypotheses that the series have a unit root can be rejected on the 99% level of confidence. KPSS test statistics are 0.117 for the sentiment index and 0.044 for market returns; hence, the null hypothesis cannot be rejected. The Schwert criterion determines the lag structure. 15 The SBIC usually favors less complex models. 10 4.3. Asymmetric responses of investor sentiment to changes in past market returns A widely discussed question in the literature is whether market participants react more severely if bad news is released.16 Putting this in other words, one should test the hypothesis that negative market returns influence sentiment more than positive returns. To implement this test into our ARCH framework, we modify model (2) by inserting an indicator variable labeled I and an interaction term of the indicator variable with lagged market returns I.rt-1.17 The indicator variable I takes the value one when the previous market return rt-1 is negative, and the value zero otherwise. Accordingly, we allow an asymmetric response of investor sentiment to past returns in the mean-equation. To explore whether asymmetric responses also exist in the variance-equation, we run three different specifications, namely the standard ARCH(2), EGARCH and TARCH models. 3 4 (3) yt 0 1rt 1 2 I 3 I rt 1 j yt j j t j t j 1 j 1 2 ARCH(2): t2 0 j t2 j j 1 2 t j 2 t j EGARCH: t vt ht ; ln( ht ) 0 j 0.5 j j 1 h t j j 1 ht0.5j 2 2 TARCH: t vt ht ; ht 0 j t2 j j Dt j t2 j j 1 j 1 Note that vt is a white noise process with variance 2v = 1, and Dt-j = 1 if t-j < 0. Table 2 shows the outcomes of model (3) with an ARCH(2) specification. To discuss asymmetric 16 Sias (1997) found that the expectations of small investors (compared to institutional investors) react more sensitively if the market environment changes. 17 Note that market returns are weakly exogenous and Granger-cause sentiment; hence, inserting lagged returns does not cause an endogeneity bias. 11 responses of investor sentiment in the conditional variance, Table 2 also reports TARCH and EGARCH model results. 18 The coefficient of the indicator variable I is not significant; however, the coefficient of the interaction term I.rt-1 is highly significant (p-value = 0.000) and has a positive sign regardless of whether we use an ARCH, EGARCH or TARCH model. Consequently, if the previous return rt-1 is negative, investor sentiment exhibits a pronounced decline. In particular, the magnitude of the impact of past returns is about three times larger in the case of negative returns compared to increasing stock prices. Based on the threshold ARCH model (see Chen, 1998), we do not find asymmetric responses in the conditional variance of forecasts triggered by past market movements. However, the EGARCH model confirms that negative shocks (measured by -j + j) have a stronger impact on conditional variance than positive shocks (measured by j + j). This confirms Chen‟s (1998) finding that investor sentiment exhibits a higher variance after negative shocks. Based on our estimates for the mean and conditional variance equations, we state that negative returns have a pronounced negative impact on investor sentiment, and that negative shocks increase the conditional variance of investor sentiment. 4.4. The impact of investor sentiment on market volatility After confirming that previous stock returns affect sentiment but not vice versa, we focus on the impact of institutional investors‟ sentiment on market volatility. To test the impact of lagged sentiment yt-1 on volatility and to capture volatility patterns, we apply a GARCH model of the following form:19 18 We use the EGARCH model developed by Nelson (1991), and the TARCH model based on Glosten, Jaganathan, and Runkle (1993). 19 Note that sentiment is a stationary variable; hence, taking the first-difference, as done by Lee, Jiang, and Indro (2002) for US data, is not required and would be a misspecification. We specify a GARCH (1,1) model after testing for the order of an ARCH (p) model using the LM test as mentioned in a previous section. Because the 12 rt 0 1 y t 1 u t (4) E u t t 0 Var u t t ut 0 1u t21 2 ut1 3 yt 1 2 2 We find that the sentiment index has a significantly negative impact on the conditional variance of market returns. Table 3 reports the results of different specifications of model (4); thus, we estimate a GARCH(1,1) model without any impact of the lagged sentiment index, a specification with the lagged sentiment index in the mean and variance equations, and a model that includes the lagged investor sentiment in the variance equation only. Accordingly, one can confirm that the sentiment has an impact on volatility – but not on returns. If the sentiment is very low, one should expect a high volatility on the market. 20 Obviously, statistical significance does not necessarily imply that the impact of sentiment on volatility is an economically important effect. To evaluate the relevance of institutional investors‟ sentiment, we determine the one-sep ahead forecasts of the conditional variance based on our GARCH(1,1) model with the lagged sentiment. Then, we express the contribution of sentiment relative to the predicted conditional variance in percentage points. In the case of overwhelming optimism (95 percentile of the sentiment indicator), market volatility drops by 12.74% on average. If institutional investors are very pessimistic (5 percentile of the sentiment indicator), volatility rises by 6.60% on average. Consequently, the impact of institutional investors‟ sentiment on market volatility has a considerable economic effect. Pessimism of institutional investors destabilizes the stock market by triggering higher market volatility. Having thus confirmed the empirical relevance of our results, do our findings also make sense theoretically? lag of the ARCH (p) model would exceed three, one can approximate this complex model by a simpler GARCH (1,1) specification (see Bollerslev, 1986). 20 In contrast to Lee, Jiang, and Indro (2002) who used an ARCH-in-mean model, we cannot confirm an impact of the sentiment on stock returns (or excess returns). However, we find an impact of sentiment on volatility, which they found for the US sentiment index. 13 Theoretical and empirical studies confirm that stocks with a higher volatility are more sensitive to shifts in investor sentiment (see Baker and Wurgler, 2004), as they exhibit a broader range of valuation, which make these stocks more vulnerable to noise trading (see DeLong, Shleifer, Summers, and Waldmann, 1990; Shleifer and Vishny, 1997). However, we investigate institutional investors‟ sentiment and focus on the opposite causal direction. A theoretical model concerning the linkage between sentiment and market volatility does not exist, but there is a broad literature on institutional trading behavior and market stability. Institutional investors can be destabilizing because of positive-feedback trading (see Lakonishok, Shleifer, and Vishny, 1992), which means that they trade less if past returns were negative, and trade more if past returns were positive. Our study shows that past returns have a positive impact on current sentiment; therefore, negative returns trigger a reduction in investor sentiment. If institutional investors are pessimistic about the future development of the stock market, they might trade less.21 What we observe is a positive-feedback trading or “trend chasing” behavior, which destabilizes exchanges as more liquidity is provided in good times and less in bad times (see Aiken, 1998). As a consequence, market volatility increases (see DeLong et al., 1990; Cutler, Poterba, and Summers, 1990). Nofsinger and Sias (1999) argued that lagged stock returns act as a common signal for positive-feedback trading, which confirms our findings that past returns affect sentiment in the sense of a positive-feedback process. 4.5. Bivariate GARCH models So far, we have focused on univariate GARCH models, and our VAR model allowed interrelations between mean-equations but did not allow volatility-spillovers and correlation between the error terms of the two-equation system. Accordingly, this section explores multivariate GARCH techniques that allow volatility-spillovers and time-varying correlation. 21 Note that we cannot observe the trading behavior of Chinese institutional investors directly. 14 We follow the bivariate BEKK model by Engle and Kroner (1995) and construct our basic model in the following manner:22 rt 3 rt j (5) VAR: μ Γ j ε yt j 1 yt j t hr ,t Vector GARCH (1,1): vec(H t ) h t hry ,t C Avec(ε t 1 ε 1 ) Bh t 1 t h y ,t BEKK parameterization: H t C C* A ε t 1ε 1 A * B H t 1 B * * * t * Based on our univariate GARCH models, we know that the conditional volatility of stock returns can be modeled by GARCH(1,1), whereas the conditional volatility of investor sentiment follows an ARCH(2) process. Accordingly, we use the standard vector GARCH(1,1) and apply the BEKK parameterization – but we also explore an extension by allowing an ARCH(2) process for investor sentiment, GARCH(1,1) for stock returns, and GARCH(1,1) for the covariance term. 23 For the sake of comparison, we also run a vector GARCH(1,1) specification without the VAR structure in the conditional mean-equation. Table 4 shows the results of our maximum-likelihood estimation. To highlight the time- varying correlation coefficients between investor sentiment and stock return, Fig.3 plots conditional correlation coefficients based on the BEKK parameterization of a VGARCH(1,1) model. 24 Fig.3 illustrates further that the conditional correlation coefficient fluctuates tremendously over time, and that on average, the correlation coefficient reaches 0.02. The 5 22 There are a huge variety of different bivariate conditional heteroscedasticity models. Bollerslev, Engle, and Wooldridge (1988) extended the univariate GARCH model to the vector GARCH – but this model does not guarantee that the variance-covariance matrix is positively semidefinite (see Lien and Luo, 1993). Hence, we rely on the BEKK model by Engle and Kroner (1995), which overcomes the problem of negative variance terms and allows modifications, as it is fairly general. 23 We can extend the vector GARCH to a VGARCH(2,1) and impose parameter restrictions. 24 Note that the two alternative specifications lead to very similar conditional correlation coefficients. 15 percentile is –0.19, and the 95 percentile is 0.21; thus, we can state that the contemporaneous correlation between the two series is of minor relevance. Nevertheless, Fig.2, the VAR in Table 1 and the extended VAR in Table 4 point to the fact that lagged returns affect investor sentiment (in the mean-equation). Based on the three bivariate GARCH specifications in Table 4, we carry out Granger causality tests to detect volatility-spillovers. Thus far, Table 3 showed that lagged investor sentiment influences the conditional variance of stock returns. Now, we focus on the alleged impact of the lagged conditional variance of investor sentiment on the conditional variance of stock returns and vice versa. However, Granger causality tests reveal that volatility-spillovers cannot be confirmed. 5. Conclusions In contrast to previous research, we analyze institutional investors‟ sentiment and reveal that a long-run relation between share prices and sentiment does not exist – but short-term dynamics matter. Granger causality tests show that previous market returns influence investors‟ sentiment – but not vice versa. Henceforth, institutional investors‟ sentiment does not predict future stock returns – but stock returns influence sentiment in the sense of a positive feedback process. Our findings confirm Brown and Cliff (2001) that discovered that the sentiment of small investors is a poor indicator of future stock returns, at least in the short run. Hence, one can state that institutional investors do not possess any superior information or above-average analytical skills to form their expectations about future market developments. An asymmetric response model emphasizes that institutional investors react by far more sensitively after the market declines. This is in line with findings for small shareholders (see Sias, 1997). Lee, Jiang, and Indro (2002) found that the sentiment of small investors and hence noise-trading increases market volatility – but also increases excess returns. This implies that the additional risk of noise trading is priced. We confirm for institutional investors that their sentiment has a tremendous impact on volatilities – but we cannot find any influence on stock 16 returns. As investor sentiment follows a positive-feedback process, we argue that Chinese institutional investors exhibit a “trend chasing” behavior, which destabilizes exchanges and hence increases market volatility (see DeLong et al., 1990; Cutler, Poterba, and Summers, 1990). As survey data on institutional investors‟ sentiment is to date only available for the Chinese market, we must carefully assess which implications can be derived for mature financial markets from our study. Institutional investors are relatively unimportant in China; however, their sentiment has a considerable impact on volatility. Henceforth, one could argue that the discovered impact of institutional investors‟ sentiment might be even stronger in mature markets. Nevertheless, Chinese institutional investors have a short history, as the first securities firms entered the market in September 1992. Due to this short period of learning, Chinese institutional investors lack experience, which might partly explain our finding that institutional investors‟ sentiment has similar effects as the sentiment of small shareholders. 17 References Aiken, B., 1998. Have institutional investors destabilized emerging markets? Contemporary Economic Policy 16, 173-184. Baker, M., Stein, J.C., 2004. Market liquidity as a sentiment indicator. Journal of Financial Markets 7, 271-299. Baker, M., Wurgler, J., 2004. Investor sentiment and the cross-section of stock returns, NBER Working Paper 10449. Barberis, N., Shleifer, A., Vishny, R., 1998. A model of investor sentiment. Journal of Financial Economics 49, 307-343. Blinder, A.S., 1991. Why are prices sticky? Preliminary results from an interview study. American Economic Review 81, 89-96. Bohl, M.T., Brzeszczynski, J., 2006. Do institutional investors destabilize stock prices? Evidence from an emerging market. Journal of International Financial Markets, Institutions & Money 16, 370-383. Bollerslev, T., 1986. Generalized autoregressive conditional heteroscedasticity. Journal of Econometrics 31, 307-327. Bollerslev, T., Engle, R.F., Wooldridge, J.M., 1988. A capital asset pricing model with time- varying covariances. Journal of Political Economy 96, 116-131. Brown, G.W., Cliff, M.T., 2001. Investor sentiment and near term stock market. Working Paper, Kenan-Flagler Business School, University of North Carolina at Chapel Hill, September 2001. Cai, F., Zheng, L., 2004. Institutional trading and stock returns. Finance Research Letters 1, 178-189. Chen, C.W.S., 1998. A Bayesian analysis of generalized threshold autoregressive models. Statistics and Probability Letters 40, 15-22. 18 Cutler, D., Poterba, J.M., Summers, L.H., 1990. Speculative dynamics and the role of feedback traders. American Economic Review Papers and Proceedings, 63-68. DeLong, J.B., Shleifer, A., Summers, L.H., Waldmann, R.J., 1990. Noise trader risk in financial markets. Journal of Political Economy 4, 703-738. Elliott, G., Rothenberg, T.J., Stock, J.H., 1996. Efficient test for an autoregressive unit root. Econometrica 64, 813-836. Engle, R.F., Kroner, K., 1995. Multivariate simultaneous generalized ARCH. Econometric Theory 11, 122-150. Engle, R., 1982. Autoregressive conditional heteroskedasticity with estimates of the variance of UK inflation. Econometrica 50, 987-1008. Glosten, L., Jagannathan, R., Runkle, D., 1993. Relationship between the expected value and the volatility of the nominal excess return on stocks. Journal of Finance 48, 1779- 1801. Gomes, J., Kogan, L., Zhang, L., 2003. Equilibrium cross section of returns. Journal of Political Economy 111, 693-732. Jones, C., Lamont, O., 2002. Short sale constraints and stock returns. Journal of Financial Economics 66, 207-239. Kelly, M., 1997. Do noise traders influence stock prices? Journal of Money, Credit, and Banking 29, 351-363. Kenneth, L.F., Statman, M., 2000. Investor sentiment and stock returns. Financial Analysts Journal, March/April, 16-23. Kwiatkowski, D., Phillips, P.C.B., Schmidt, P., Shin, Y., 1992. Testing the null hypothesis of stationarity against the alternative of a unit root. Journal of Econometrics 54, 159–178. Lakonishok, J., Shleifer, A., Vishny, R.W., 1992. The impact of institutional trading on stock prices. Journal of Financial Economics 32, 23-43. 19 Lee, W.Y., Jiang, C.X., Indro, D.C., 2002. Stock market volatility, excess returns, and the role of investor sentiment. Journal of Banking and Finance 26, 2277-2299. Lee, C.M.C., Shleifer, A., Thaler, R.D., 1991. Investor sentiment and the closed-end fund puzzle. Journal of Finance 46, 75-109. Lien, D., Luo, X., 1993. Estimating multiperiod hedge ratios in cointegrated markets. Journal of Futures Markets 13, 909-920. Neal, R., Wheatley, S.M., 1998. Do measures of investor sentiment predict return? Journal of Financial and Quantitative Analysis 33, 1775-1798. Nelson, D.B., 1991. Conditional heteroskedasticity in asset returns: a new approach, Econometrica 59, 347-370. Nofsinger, J.R., Sias, R.W., 1999. Herding and feedback trading by institutional investors. Journal of Finance 54, 2263-2295. Shleifer, A., Vishny, R.W., 1997. The limits of arbitrage. Journal of Finance 52, 35-55. Sias, R.W., 1997. The sensitivity of individual and institutional investors‟ expectations to changing market conditions: evidence from closed-end funds. Review of Quantitative Finance and Accounting 8, 245-269. Siegel, J.J., 1992. Equity risk premia, corporate profit forecasts, and investor sentiment around the stock crash of October 1987. Journal of Business 65, 557-570.

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