Bootstrap method

A computer-intensive re-sampling method, introduced in statistics by B. Efron in 1979 ([a3]) for estimating the variability of statistical quantities and for setting confidence regions. The name ‘bootstrap’ refers to the analogy of pulling oneself up by one’s own bootstraps. Efron’s bootstrap is to re-sample the data — given observations $ X_{1},\ldots,X_{n} $, artificial bootstrap samples are drawn with replacement from $ X_{1},\ldots,X_{n} $, putting an equal probability mass of $ \dfrac{1}{n} $ on $ X_{i} $ for each $ i \in \{ 1,\ldots,n \} $. For example, with a sample size of $ n = 5 $ and distinct observations $ X_{1},X_{2},X_{3},X_{4},X_{5} $, one might obtain $ X_{3},X_{3},X_{1},X_{5},X_{4} $ as a bootstrap sample. In fact, there are $ 126 $ distinct bootstrap samples in this case.

A more rigorous description of Efron’s non-parametric bootstrap in a simple setting is as follows. Suppose that $ (X_{1},\ldots,X_{n}) $ is a random sample of size $ n $ drawn from a population whose underlying probability space is $ (\Omega,\mathscr{S},\mathsf{P}) $, i.e., the $ X_{i} $’s are independent and identically-distributed random variables on $ (\Omega,\mathscr{S},\mathsf{P}) $. Let $ F $ denote the common cumulative distribution function (cdf) of the $ X_{i} $’s, which we assume to be unknown. Let $ \Theta $ be some pre-specified (non-linear) functional on the space of cdf’s; the parameter that we want to estimate is $ \theta \stackrel{\text{df}}{=} \Theta(F) $. Let $ T: \mathbf{R}^{n} \to \mathbf{R} $ be a statistical estimator for $ \theta $ based on $ (X_{1},\ldots,X_{n}) $ (cf. also statistical estimation), and let $ T_{n} \stackrel{\text{df}}{=} T(X_{1},\ldots,X_{n}) $. The object of interest is then the cdf $ G_{n} $ of the random variable $ \sqrt{n} (T_{n} - \theta) $ on $ (\Omega,\mathscr{S},\mathsf{P}) $ defined by $$ \forall x \in \mathbf{R}: \qquad {G_{n}}(x) \stackrel{\text{df}}{=} \mathsf{P}(\{ \omega \in \Omega \mid \sqrt{n} [{T_{n}}(\omega) - \theta] \leq x \}). $$ This is the cdf of $ T_{n} $ properly normalized. The scaling factor $ \sqrt{n} $ is a classical one, while the centering of $ T_{n} $ is by the parameter $ \theta $.

Efron’s non-parametric bootstrap estimator of $ G_{n} $, which we denote by $ G_{n}^{\ast} $, is now defined by $$ \forall x \in \mathbf{R}, ~ \forall \omega \in \Omega: \qquad {G_{n}^{\ast}}(x;\omega) \stackrel{\text{df}}{=} [{\mathsf{P}_{n}^{\ast}}(\omega)](\{ \mathbf{a} \in \{ {X_{1}}(\omega),\ldots,{X_{n}}(\omega) \}^{n} \mid \sqrt{n} [[{T_{n}^{\ast}}(\omega)](\mathbf{a}) - {\theta_{n}}(\omega)] \leq x \}). $$ Here, $ {T_{n}^{\ast}}(\omega) \stackrel{\text{df}}{=} T({X_{1}^{\ast}}(\omega),\ldots,{X_{n}^{\ast}}(\omega)) $, where we have the following:

• $ ({X_{1}^{\ast}}(\omega),\ldots,{X_{n}^{\ast}}(\omega)) $ is a random sample (the bootstrap sample) drawn from the set $ \{ {X_{1}}(\omega),\ldots,{X_{n}}(\omega) \}^{n} $.
• For each $ i \in \{ 1,\ldots,n \} $, the random variable $ {X_{i}^{\ast}}(\omega) $ is defined by $ [{X_{i}^{\ast}}(\omega)](\mathbf{a}) \stackrel{\text{df}}{=} \mathbf{a}(i) $ for all $ \mathbf{a} \in \{ {X_{1}}(\omega),\ldots,{X_{n}}(\omega) \}^{n} $.

The cdf of the $ {X_{i}^{\ast}}(\omega) $’s is designated to be $ {\hat{F}_{n}}(\bullet;\omega) $, where $ \hat{F}_{n} $ denotes the empirical cdf associated with the sample $ (X_{1},\ldots,X_{n}) $. Note that $ \hat{F}_{n} $ (which is a random step function) puts a probability mass of $ \dfrac{1}{n} $ on $ X_{i} $ for each $ i \in \{ 1,\ldots,n \} $, and it is sometimes referred to as the re-sampling distribution. Finally, $ {\mathsf{P}_{n}^{\ast}}(\omega) $ denotes the probability measure on $ \{ {X_{1}}(\omega),\ldots,{X_{n}}(\omega) \}^{n} $ that corresponds to $ {\hat{F}_{n}}(\bullet;\omega) $, and $ {\theta_{n}}(\omega) \stackrel{\text{df}}{=} \Theta \! \left( {\hat{F}_{n}}(\bullet;\omega) \right) $.

Obviously, given an observed sample $ ({X_{1}}(\omega),\ldots,{X_{n}}(\omega)) $, the cdf $ {\hat{F}_{n}}(\bullet;\omega) $ is completely known and hence, $ {G_{n}^{\ast}}(\bullet;\omega) $ is also completely known (at least in principle). One may view $ G_{n}^{\ast} $ as the empirical counterpart in the ‘bootstrap world’ to $ G_{n} $ in the ‘real world’. In practice, an exact computation of $ G_{n}^{\ast} $ is usually impractical (for an observed sample $ ({X_{1}}(\omega),\ldots,{X_{n}}(\omega)) $ consisting of $ n $ distinct numbers, there are $ \displaystyle \binom{2 n - 1}{n} $ distinct bootstrap samples), but $ G_{n}^{\ast} $ can be approximated by means of Monte-Carlo simulation. Efficient bootstrap simulation is discussed, for example, in [a2] and [a10].

When does Efron’s bootstrap work? The consistency of the bootstrap approximation $ G_{n}^{\ast} $ viewed as an estimate of $ G_{n} $, i.e., the requirement that $$ \lim_{n \to \infty} \left[ \sup_{x \in \mathbf{R}} |{G_{n}^{\ast}}(x;\omega) - {G_{n}}(x)| \right] = 0, \quad \mathsf{P}\text{-a.s. in} ~ \omega, $$ is generally viewed as an absolute prerequisite for Efron’s bootstrap to work in the problem at hand. Of course, bootstrap consistency is only a first-order asymptotic result, and the error committed when $ G_{n} $ is estimated by $ G_{n}^{\ast} $ may still be quite large in finite samples. Second-order asymptotics (cf. Edgeworth series) enables one to investigate the speed at which $ \displaystyle \sup_{x \in \mathbf{R}} |{G_{n}^{\ast}}(x;\omega) - {G_{n}}(x)| $ approaches $ 0 $, and also to identify cases where the rate of convergence is faster than $ \dfrac{1}{\sqrt{n}} $ — the classical Berry-Esseen-type rate for the normal approximation. An example in which the bootstrap possesses the beneficial property of being more accurate than the traditional normal approximation is the Student $ t $-statistic and, more generally, Studentized statistics. For this reason, the use of bootstrapped Studentized statistics for setting confidence intervals is strongly advocated in a number of important problems. A general reference is [a7].

When does the bootstrap fail? It has been proved in [a1] that in the case of the mean, Efron’s bootstrap fails when $ F $ is the domain of attraction of an $ \alpha $-stable law with $ 0 < \alpha < 2 $. However, by re-sampling from $ \hat{F}_{n} $ with a (smaller) re-sample size $ m(n) $ that satisfies $ m(n) \to \infty $ and $ \dfrac{m(n)}{n} \to 0 $ as $ n \to \infty $, it can be shown that the (modified) bootstrap works. More generally, in recent years, the importance of a proper choice of the re-sampling distribution has become clear (see [a5], [a9] and [a10]).

The bootstrap can be an effective tool in many problems of statistical inference, for example, the construction of a confidence band in non-parametric regression, testing for the number of modes of a density, or the calibration of confidence bounds (see [a2], [a4] and [a8]). Re-sampling methods for dependent data, such as the block bootstrap, is another important topic of recent research (see [a2] and [a6]).

References

[a1]	K.B. Athreya, “Bootstrap of the mean in the infinite variance case”, Ann. Statist., 15 (1987), pp. 724–731.
[a2]	A.C. Davison, D.V. Hinkley, “Bootstrap methods and their application”, Cambridge Univ. Press (1997).
[a3]	B. Efron, “Bootstrap methods: another look at the jackknife”, Ann. Statist., 7 (1979), pp. 1–26.
[a4]	B. Efron, R.J. Tibshirani, “An introduction to the bootstrap”, Chapman & Hall (1993).
[a5]	E. Giné, “Lectures on some aspects of the bootstrap”, P. Bernard (ed.), Ecole d'Eté de Probab. Saint Flour XXVI-1996, Lecture Notes Math., 1665, Springer (1997).
[a6]	F. Götze, H.R. Künsch, “Second order correctness of the blockwise bootstrap for stationary observations”, Ann. Statist., 24 (1996), pp. 1914–1933.
[a7]	P. Hall, “The bootstrap and Edgeworth expansion”, Springer (1992).
[a8]	E. Mammen, “When does bootstrap work? Asymptotic results and simulations”, Lecture Notes Statist., 77, Springer (1992).
[a9]	H. Putter, W.R. van Zwet, “Resampling: consistency of substitution estimators”, Ann. Statist., 24 (1996), pp. 2297–2318.
[a10]	J. Shao, D. Tu, “The jackknife and bootstrap”, Springer (1995).