Abstract
Academics and practitioners have frequently debated the relationship between market capitalization and expected return. We apply the Markowitz efficient frontier approach to develop a portfolio performance measure that compares the return of a portfolio to its optimal return, using data from the UK stock market over the period 1985–2012. Our results show that there is a negative relationship between portfolio size and portfolio return during the period under study. When comparing actual portfolio return with achievable return for the same level of risk, we find that as the portfolio size expands, underperformance of the portfolio increases, i.e. the larger the portfolio size, the greater the underperformance. This indicates that Markowitz efficiency is difficult to achieve, particularly in large portfolios. Changing model parameters leads to alternative efficient frontiers that impact upon the measurement of performance. However, the use of alternative efficient frontiers does not affect our result of the size effect on the relative performance of portfolios. Our study shows that the size effect is present over the full period. Our findings also suggest that the excess returns found in small portfolios are likely to be associated with higher levels of diversifiable risk in comparison with larger portfolios. Furthermore, in contrast to other studies, we find no evidence to support the size reversal effect in the data.
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Notes
Earlier studies by Reinganum (1981), Banz (1981), and Basu (1983) do not support the implication that the size anomaly is caused by capital market inefficiency; they argue that the size anomaly results from the misspecification of the CAPM as it omits other risk factors in pricing common stocks. These studies have shown inconsistent results. Reinganum (1981) shows that the smallest firms in a given E/P quintile systematically outperform the high market value firms in that quintile. There is a significant size effect irrespective of controlling a portfolio’s E/P ratio. However, when controlling the market values of portfolios, the E/P effect is not found. Consequently, the risk factors missing from the CAPM specification appear to be more closely associated with firm size than E/P ratios. Further, Reinganum (1981) concludes that the size effect subsumes the E/P effect. In contrast, Basu (1983) finds a significant E/P effect is present when controlling the market values in a given E/P portfolio. That is, portfolios with high E/P ratios produce higher risk-adjusted returns than those with low E/P ratios. However, the abnormal returns on small size portfolios are not significantly and statistically different from zero when E/P ratios and market betas are controlled. Accordingly, Basu (1983) does not support Reinganum’s results.
The first derivative of \( f (\upsigma_{\text{p}}^{ 2} ) \) is positive and the second derivative is negative, implying that the efficient frontier curve is monotonously concave down in the return-risk space.
The first derivative of \( f (\upsigma_{\text{p}}^{ 2} ) \) is \( f^{\prime } (\upsigma_{\text{p}}^{ 2} ) = \frac{1}{2}\frac{1}{c}\left[ {cd\left( {\upsigma_{\text{p}}^{ 2} - \frac{1}{c}} \right)} \right]^{{ - \frac{1}{2}}} cd > 0 \).
Since \( \frac{ 1}{\text{c}} \) is the variance of the GMV portfolio indicating that every portfolio on the mean–variance frontier has a higher variance than the GMV portfolio, \( \upsigma_{\text{p}}^{ 2} - \frac{1}{c} \) must be positive. Further, since both c and d are greater than zero, \( f'\left( {\upsigma_{\text{p}}^{ 2} } \right) \) is positive.
The second derivative is \( f^{\prime \prime } (\upsigma_{\text{p}}^{ 2} ) = - \frac{1}{4}\frac{ 1}{\text{c}}{\text{cd}}\left( {\upsigma_{\text{p}}^{ 2} - \frac{ 1}{\text{c}}} \right)^{{ - \frac{3}{2}}} \)and it is negative. The quadratic function explains that investors are risk averse when investing in risky assets. Investors will demand a higher return on investments that have a higher level of risk. Since the function \( f (\upsigma_{\text{p}}^{ 2} ) \) is concave down, this means that risk-averse investors will not invest in risky assets unless they are compensated for bearing such increased risk.
With a set of target expected return and variance combinations that minimizes the portfolio variance for a given level of return, Eq. (7) can be used to sketch a minimum-variance frontier. By definition, portfolios on the efficient frontier have a positive relationship between risk and return. Since the slope of the efficient frontier is positive, the minimum-variance frontier is different from the efficient frontier, which is divided at the GMV portfolio on the top half of the minimum-variance frontier.
The Markowitz deficit ratio compares a portfolio return relative to its return on the efficient frontier. Since the efficient frontier is on the top half of the mean–variance frontier, when a portfolio has a return less than GMV, it means that the MD is greater than 100 %. This infers that the performance of the portfolio is below GMV expected return.
The result of the portfolio performance comparison for two subperiods based on the Sharpe ratio is likely to be different from the MD ratio as the Sharp measure is affected by the levels of the risk-free rate.
In mean–variance space for excess returns, the slope of the upper mean-variance frontier is \( \sqrt {\text{d/c}} \) and the lower mean-variance frontier is \( - \sqrt {\text{d/c}} \). The slope represents the best return-to-risk ratio for risky assets on the efficient frontier. The slope and GMV are used to construct the efficient frontier (see Jorion 2003).
The expected return on the GMV is expected to be greater than the risk-free rate. The choice of the GMV return is at authors’ discretion but it is affected by the prevailing risk-free rate. While the average monthly 3-month UK Treasury bill was 0.5 % over the period 1985–2012, it was 0.3 % over the period 2000–2012.
Kim and Burnie (2002) compare portfolio performance for small firms within the expansion and contraction periods and they show that small firms perform significantly better in the expansion periods than they do in the contraction periods.
The least squares estimator is an alternative approach to estimate bootstrapped standard errors. The bootstrap estimator of the asymptotic covariance matrix \( {{\Upsigma}}_{\text{b}} \) is \( {\text{R}}^{ - 1} \sum\nolimits_{\text{i = 1}}^{\text{R}} {\left( {{\text{X}}{}_{\text{i}} - \overline{\text{X}} } \right)\left( {{\text{X}}{}_{\text{i}} - \overline{\text{X}} } \right)^{\prime } } \), where R is the resampling times with replacement; \( {\text{X}}{}_{\text{i}} \) is the ith least squares estimate based on a sample of n observations, drawn with replacement, from the original data set; \( {\bar{\text{X}}} \) is the least squares estimator of the coefficients. The bootstrapped estimated standard errors are the square roots of the diagonal elements of the covariance matrix for \( {{\Upsigma}}_{\text{b}} \) (see Greene 2003).
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Acknowledgments
The authors would like to thank Prof. Cheng-Few Lee, Editor in Chief, Review of Quantitative Finance and Accounting and two anonymous reviewers for their constructive comments and suggestions that have helped to improve the quality of the paper.
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Appendix
Appendix
We address the issue concerning the impact on our results if our sample had not required survival for the period. Since this paper collected stocks of the FTSE All-Share Index at June 2012 starting from January 1985, the portfolio returns in Panels A and B in Table 2 are based on existing stocks that continue to survive from the testing periods, January 1985–June 2012 and January 2000–June 2012. To investigate whether the results differ if the sample had not required survival for the full period starting from 1985, we include four other subperiods for reporting the size effect: January 1990–June 2012, January 1995–June 2012, January 2000–June 2012, and January 2005–June 2012. In doing so, we can increase our sample size to further examine the size effect over different periods. Results in Table 9 show that the smallest portfolio persistently outperforms the largest portfolio. For example, over the period between 1995 and 2012 the monthly difference in SML (small minus large portfolio) is 0.54 %, resulting in a gain of 6.5 % per annum. Table 9 also exhibits that the monthly standard deviation of each size portfolio is relatively stable between 4.87 % (portfolio 5 over the period 1995–2012 in Panel A) and 6.89 % (portfolio 7 over the period 2005–2012 in Panel B).
In Table 10 we include 612 stocks on the FTSE All-Share Index by forming 5 and 10 portfolios to calculate their portfolio returns respectively. We split the sample into quintiles to form 5 portfolios and also divide the sample into deciles to form 10 portfolios. Since the maximum number of stocks in our sample is 612 stocks in 2012, many of these stocks will not be present at the start of the period under review. The formation of size portfolios can only be based on the ascending order of market capitalization of stocks at June 2012 rather than at the beginning of 1985. Although we are aware that if investors perceive the size effect they should form their size portfolios based on the forward-looking portfolio formation (i.e. at the beginning of the time when they purchase stocks), the backward-looking portfolio formation helps to explain the possibility of survivorship bias. The results shown in Table 10 indicate the final portfolio position of the available stocks in 2012. The return bias is not comparable with Table 2 in which portfolios are formed at the beginning of observational period.
Table 10 shows that the smallest size portfolio tends to have a lower return than larger size portfolios. Although at first sight these results appear to contradict our findings of the size effect, after further examination these figures clearly show that the size effect is present in the data. Since we rank portfolios based on the end of the observational period at June 2012, several small stocks when purchased at the beginning of the observational research period yielded higher returns and increased their market capitalizations, gradually expanding in size. Accordingly, the remaining stocks in the smallest size portfolio are the least successful companies, confirming the size effect in the UK stock market.
Table 11 reports the increase in market capitalization for each size portfolio from the start of each respective period of portfolio formation to June 2012 showing only companies that have survived for the entire periods of observation. The average size of the smallest portfolio in Panel A is 1,027 million pounds and the market capitalization growth rate of the smallest portfolio is 20 % per year. In contrast, all other portfolios have market capitalization growth rates between 8 and 11 % in Panel A. When investigating the short observational period 2000–2012, the smallest portfolio also exhibits the highest growth rate among all size portfolios. Although the smallest size portfolio has a high growth rate, the growth rate only counts for those stocks which continue to exist in the sample period. This growth rate for the smallest portfolio may be optimistic as the failing firms removed from the index are more likely to be in the smallest group and have significantly lower returns than surviving firms. To summarize the results in the Appendix, consideration of additional subperiods and regrouping the number of stocks in portfolios strengthens the results.
To explore the survivorship bias, Kothari et al. (1995) collect two sets of stock return data from COMPUSTAT data and CRSP (Centre for Research in Securities Prices) data and they assume that the CRSP sample does not suffer from a survivorship bias problem. Kothari et al. (1995) find that the returns for small firms are 9–10 % points higher on COMPUSTAT than returns for CRSP—COMPUSTAT small firms. Since our sample to estimate portfolio returns is based on available stocks at the end of 2012, the survivorship bias in our sample can be regarded as the difference between portfolio return on the existing stocks over the entire observational period and portfolio return on the existing stocks over the different subperiods. Thus, it can be argued that portfolio return on available stocks for the full period is likely to be higher than the portfolio return on available stocks for a subperiod.
where \( {\text{SB}}_{\text{j}}^{ 1 9 8 5} \) indicates survivorship bias of portfolio j whose portfolio return is based on available stocks since 1985. \( {\bar{\text{r}}}_{{{\text{j}},1985}} \) (\( {\bar{\text{r}}}_{{{\text{j}},2000}} \)) is portfolio j return for available stocks since 1985 (2000) as shown in Table 2. The estimation of the survivorship bias based on the smallest size portfolio can be derived between Tables 2 and 9 (\( {\bar{\text{r}}}_{{{\text{j}},2005}} \) is portfolio j return for available stocks since 2005 as shown in Table 9). The monthly portfolio bias for \( {\text{SB}}_{\text{j}}^{ 1 9 8 5} \) on the smallest portfolio is 0.31 %, resulting in a bias of 3.7 % per annum. Further, the monthly portfolio bias for \( {\text{SB}}_{\text{j}}^{ 2 0 0 0} \) on the smallest portfolio is 0.38 %, resulting in a bias of 4.5 % per annum. The average survivorship bias of the smallest size portfolio is 4.1 % which is much lower than the 9–10 % points in Kothari et al. (1995). The resulting difference of our estimation is that we only consider the available stocks at the start of each period under investigation.
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Hwang, T., Gao, S. & Owen, H. Markowitz efficiency and size effect: evidence from the UK stock market. Rev Quant Finan Acc 43, 721–750 (2014). https://doi.org/10.1007/s11156-013-0390-8
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DOI: https://doi.org/10.1007/s11156-013-0390-8