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In Exercise \(10.9 .5\), the influence function of the variance functional was derived directly. Assuming that the mean of \(X\) is 0 , note that the variance functional, \(V\left(F_{X}\right)\), also solves the equation $$ 0=\int_{-\infty}^{\infty}\left[t^{2}-V\left(F_{X}\right)\right] f_{X}(t) d t $$ (a) Determine the natural estimator of the variance by writing the defining equation at the empirical cdf \(F_{n}(t)\), for \(X_{1}-\bar{X}, \ldots, X_{n}-\bar{X}\) iid with \(\operatorname{cdf} F_{X}(t)\) and solving for \(V\left(F_{n}\right)\). (b) As in Exercise \(10.9 .6\), write the defining equation for the variance functional at the contaminated \(\operatorname{cdf} F_{x, \epsilon}(t)\). (c) Then derive the influence function by implicit differentiation of the defining equation in part (b).

Short Answer

Expert verified
Natural estimator of the variance: \(V\left(F_{n}\right)=\frac{1}{n}\sum_{i=1}^{n}(x_{i}-\bar{x})^{2}\). The influence function is given by: \[IF(x, F_X) = \frac{(x-\bar{x})^2 - V(F_X)}{2}.\]

Step by step solution

01

Set Up Defining Equation at Empirical cdf

Accordingly, replace \(F_X(t)\) and \(f_X(t)\) in the integral equation by \(F_n(t)\) and the sum of Dirac functions, respectively. Therefore, we have: \[0=\sum_{i=1}^{n}\left[(x_{i}-\bar{x})^{2}-V\left(F_{n}\right)\right] / n. \]Rearrange to solve for \(V\left(F_{n}\right)\), we get:\[V\left(F_{n}\right)=\frac{1}{n}\sum_{i=1}^{n}(x_{i}-\bar{x})^{2}\].
02

Write Defining Equation for the Contaminated cdf

The definition of \(F_{X, \epsilon}\) as a contaminated cdf and the corresponding pdf can be used to modify the integral equation as follows:\[0=\int[t^2-V(F_{X,\epsilon})]f_{X,\epsilon}(t) dt\]where:\[F_{X,\epsilon}(t) =(1-\epsilon)F_X(t) + \epsilon F_X^D(t)\] and\[f_{X,\epsilon}(t) = (1-\epsilon)f_X(t) + \epsilon f_X^D(t)\].
03

Derive the Influence Function

To derive the Influence Function (IF) take the derivative of the equation in Step 2 with respect to \(\epsilon\), and then let the limit as \(\epsilon \rightarrow 0\). Solve for the IF, we yield:\[\frac{dF_{X,\epsilon}}{d\epsilon} = \lim_{\epsilon\rightarrow 0} \frac{F_{X,\epsilon} - F_{X}}{\epsilon}\]Then after some mathematical manipulations:\[IF(x, F_X) = \lim_{\epsilon\rightarrow 0} \frac{F_{X,\epsilon}- F_{X}}{\epsilon} = \frac{(x-\bar{x})^2 - V(F_X)}{2}.\]

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Most popular questions from this chapter

Consider the location Model (10.3.35). Assume that the pdf of the random errors, \(f(x)\), is symmetric about \(0 .\) Let \(\widehat{\theta}\) be a location estimator of \(\theta\). Assume that \(E\left(\widehat{\theta}^{4}\right)\) exists. (a) Show that \(\widehat{\theta}\) is an unbiased estimator of \(\theta\). Hint: Assume without loss of generality that \(\theta=0 ;\) start with \(E(\hat{\theta})=\) \(E\left[\widehat{\theta}\left(X_{1}, \ldots, X_{n}\right)\right]\); and use the fact that \(X_{i}\) is symmetrically distributed about \(0 .\) (b) As in Section \(10.3 .4\), suppose we generate \(n_{s}\) independent samples of size \(n\) from the pdf \(f(x)\) which is symmetric about \(0 .\) For the \(i\) th sample, let \(\widehat{\theta}_{i}\) be the estimate of \(\theta\). Show that \(n_{s}^{-1} \sum_{i=1}^{n_{x}} \widehat{\theta}_{i}^{2} \rightarrow V(\hat{\theta})\), in probability.

Prove that a pdf (or pmf) \(f(x)\) is symmetric about 0 if and only if its mgf is symmetric about 0, provided the mgf exists.

Consider the hypotheses (10.4.4). Suppose we select the score function \(\varphi(u)\) and the corresponding test based on \(W_{\varphi} .\) Suppose we want to determine the sample size \(n=n_{1}+n_{2}\) for this test of significance level \(\alpha\) to detect the alternative \(\Delta^{*}\) with approximate power \(\gamma^{*}\). Assuming that the sample sizes \(n_{1}\) and \(n_{2}\) are the same, show that $$ n \approx\left(\frac{\left(z_{\alpha}-z_{\gamma^{*}}\right) 2 \tau_{\varphi}}{\Delta^{*}}\right)^{2} $$

Suppose \(X\) is a random variable with mean 0 and variance \(\sigma^{2}\). Recall that the function \(F_{x, \epsilon}(t)\) is the cdf of the random variable \(U=I_{1-e} X+\left[1-I_{1-e}\right] W\), where \(X, I_{1-\epsilon}\), and \(W\) are independent random variables, \(X\) has cdf \(F_{X}(t), \underline{W}\) has cdf \(\Delta_{x}(t)\), and \(I_{1-\epsilon}\) has a binomial \((1,1-\epsilon)\) distribution. Define the functional \(\operatorname{Var}\left(F_{X}\right)=\operatorname{Var}(X)=\sigma^{2}\). Note that the functional at the contaminated \(\operatorname{cdf} F_{x, c}(t)\) has the variance of the random variable \(U=I_{1-e} X+\left[1-I_{1-\epsilon}\right] W\). To derive the influence function of the variance, perform the following steps: (a) Show that \(E(U)=\epsilon x\). (b) Show that \(\operatorname{Var}(U)=(1-\epsilon) \sigma^{2}+\epsilon x^{2}-\epsilon^{2} x^{2}\) (c) Obtain the partial derivative of the right side of this equation with respect to \(\epsilon\). This is the influence function. Hint: Because \(I_{1-e}\) is a Bernoulli random variable, \(I_{1-\epsilon}^{2}=I_{1-e} .\) Why?

Consider the location model as defined in expression (10.9.1). Let $$ \widehat{\theta}=\operatorname{Argmin}_{\theta}\|\mathbf{X}-\theta \mathbf{1}\|_{\mathrm{LS}}^{2} $$ where \(\|\cdot\|_{\mathrm{LS}}^{2}\) is the square of the Euclidean norm. Show that \(\widehat{\theta}=\bar{x}\).

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