Electronic copy available at: https://ssrn.com/abstract=2000901
pairs for which the null hypothesis is correct) and p–values that are close to zero (when the null hypothesis
is false). This key insight is provided by Storey (2002) and it is used by Barras et al. (2010) to estimate the
proportion of funds that perform equally well in the same manner as a passive investor in style indices.
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The next step is then to attribute the remaining segment of the peer group to funds that significantly
underperform and those that outperform. Thus, for each fund, we obtain an equal–performance ratio, an
outperformance ratio, and an underperformance ratio. The proposed equal–performance ratio is robust to
false discoveries, and unless the fund performance differs significantly from its peers, it will tend toward
100%. When the return data are sufficiently informative about differences in performance, the outperfor-
mance ratio is suitable for classifying funds into top–performing funds. The estimators rely on pairwise
tests to calculate the p–values, thus they can use the longest common time series span for each pair in an op-
timal manner, and parallel computing can be employed to calculate these p–values in a numerically efficient
way.
2.4. The equal–performance ratio
A crucial feature of the proposed peer performance ratios is that they exploit the difference in the
distribution of the p–values bp
i−j
when ∆
i−j
= 0 versus ∆
i−j
6= 0.
If the test is sufficiently powerful, a threshold value λ
i
exists such that almost surely the p–value of the
two–sided equal–performance test is less than λ
i
if the two funds have different performances:
(A1) : P[bp
i−j
< λ
i
| ∆
i−j
6= 0] = 1 . (7)
The validity of this assumption depends on the magnitude of ∆
i−j
, the test statistic itself, the calculation of
its p–value (e.g., asymptotic versus bootstrap methods), and the sample size.
In the case of equal performance, and provided that the estimated
b
F
i−j
coincides with the true F
i−j
used to calculate the p–value, the p–value is uniformly distributed for a given pair (i, j). This implies that
the probability of bp
i−j
exceeding λ
i
when ∆
i−j
= 0 is 1 − λ
i
:
(A2) : P[bp
i−j
≥ λ
i
| ∆
i−j
= 0] = 1 − λ
i
. (8)
In practice, the cumulative distribution function F
i−j
is not fully known and the calculation of the p–values
requires parameter estimates. Asymptotically, the p–value is uniformly distributed if consistent estimators
are used (Rosenblatt, 1952), whereas in finite samples, assumption (A2) is only approximately satisfied.
A key result related to the definition of the proposed equal–performance ratio is that under (A1) and
(A2), the expected number of p–values exceeding λ
i
is (1 − λ
i
)n
0
i
, with n
0
i
the number of peer funds that
perform equally well as the focal fund:
E
X
j6=i
I{bp
i−j
≥ λ
i
}
= E
X
j6=i
∆
i−j
=0
I{bp
i−j
≥ λ
i
}
| {z }
= (1−λ
i
) n
0
i
+ E
X
j6=i
∆
i−j
6=0
I{bp
i−j
≥ λ
i
}
| {z }
= 0
= (1 − λ
i
) n
0
i
.
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Instead of estimating the market–wide equal–performance ratio, as in Barras et al. (2010), we estimate the individual equal–
performance ratio for each fund, which broadens the application scope. The aggregate equal–performance ratio allows us to answer
general economic questions such as the usefulness of actively managed funds, but our proposed individual equal–performance ratio
can be used directly by investors to evaluate the performance of a specific fund.
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