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Suppose \(\boldsymbol{Y}\) is an \(n \times 1\) random vector, \(\boldsymbol{X}\) is an \(n \times p\) matrix of known constants of rank \(p\), and \(\beta\) is a \(p \times 1\) vector of regression coefficients. Let \(\boldsymbol{Y}\) have a \(N\left(\boldsymbol{X} \boldsymbol{\beta}, \sigma^{2} \boldsymbol{I}\right)\) distribution. Discuss the joint pdf of \(\hat{\boldsymbol{\beta}}=\left(\boldsymbol{X}^{\prime} \boldsymbol{X}\right)^{-1} \boldsymbol{X}^{\prime} \boldsymbol{Y}\) and \(\boldsymbol{Y}^{\prime}\left[\boldsymbol{I}-\boldsymbol{X}\left(\boldsymbol{X}^{\prime} \boldsymbol{X}\right)^{-1} \boldsymbol{X}^{\prime}\right] \boldsymbol{Y} / \sigma^{2}\)

Short Answer

Expert verified
The joint pdf of the regression coefficient estimator \(\hat{\boldsymbol{\beta}}\) and the estimator for residual sum of squares \(\boldsymbol{Y}^{\prime}\left[\boldsymbol{I}-\boldsymbol{X}\left(\boldsymbol{X}^{\prime} \boldsymbol{X}\right)^{-1} \boldsymbol{X}^{\prime}\right] \boldsymbol{Y} / \sigma^{2}\) is the product of their respective distributions. \(\hat{\boldsymbol{\beta}}\) follows a multivariate normal distribution while the residual sum of squares follows a chi-square distribution with n-p degrees of freedom.

Step by step solution

01

Joint Distribution

The distribution of the vector \(\boldsymbol{Y}\) is given by the following equation: \(\boldsymbol{Y} \sim N\left(\boldsymbol{X} \boldsymbol{\beta}, \sigma^{2} \boldsymbol{I}\right)\). The normal equations of the linear regression model are \(\boldsymbol{X}^{\prime}\boldsymbol{X}\hat{\boldsymbol{\beta}} = \boldsymbol{X}^{\prime} \boldsymbol{Y}\), which is the basis of finding \(\hat{\boldsymbol{\beta}}\).
02

Estimator for Regression Coefficients

Solving the normal equations for \(\hat{\boldsymbol{\beta}}\), we find that \(\hat{\boldsymbol{\beta}} = \left(\boldsymbol{X}^{\prime} \boldsymbol{X}\right)^{-1}\boldsymbol{X}^{\prime} \boldsymbol{Y}\). Therefore, the distribution for \(\hat{\boldsymbol{\beta}}\) is as follows: \(\hat{\boldsymbol{\beta}} \sim N\left(\boldsymbol{\beta}, \sigma^{2}\left(\boldsymbol{X}^{\prime} \boldsymbol{X}\right)^{-1}\right)\)
03

Estimator for Residual Sum of Squares

The quantity \(\boldsymbol{Y}^{\prime}\left[\boldsymbol{I}-\boldsymbol{X}\left(\boldsymbol{X}^{\prime} \boldsymbol{X}\right)^{-1} \boldsymbol{X}^{\prime}\right] \boldsymbol{Y} / \sigma^{2}\) represents the residual sum of squares. It can be shown that, with degrees of freedom n-p, it follows a chi-square distribution, represented as: \(\boldsymbol{Y}^{\prime}\left[\boldsymbol{I}-\boldsymbol{X}\left(\boldsymbol{X}^{\prime} \boldsymbol{X}\right)^{-1} \boldsymbol{X}^{\prime}\right] \boldsymbol{Y} / \sigma^{2} \sim \chi^{2}_{(n-p)}\)
04

Joint pdf

We need to identify the joint pdf of \(\hat{\boldsymbol{\beta}}\) and \(\boldsymbol{Y}^{\prime}\left[\boldsymbol{I}-\boldsymbol{X}\left(\boldsymbol{X}^{\prime} \boldsymbol{X}\right)^{-1} \boldsymbol{X}^{\prime}\right] \boldsymbol{Y} / \sigma^{2}\). As these quantities are independent and derived from a normally distributed random vector, the joint pdf would be the product of their individual pdfs, which are normal and chi-square respectively.

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Most popular questions from this chapter

Show that \(R=\frac{\sum_{1}^{n}\left(X_{i}-\bar{X}\right)\left(Y_{i}-\bar{Y}\right)}{\sqrt{\sum_{1}^{n}\left(X_{i}-\bar{X}\right)^{2} \sum_{1}^{n}\left(Y_{i}-Y\right)^{2}}}=\frac{\sum_{1}^{n} X_{i} Y_{i}-n \overline{X Y}}{\sqrt{\left(\sum_{1}^{n} X_{i}^{2}-n \bar{X}^{2}\right)\left(\sum_{1}^{n} Y_{i}^{2}-n \bar{Y}^{2}\right)}}\)

Using the notation of this section, assume that the means satisfy the condition that \(\mu=\mu_{1}+(b-1) d=\mu_{2}-d=\mu_{3}-d=\cdots=\mu_{b}-d .\) That is, the last \(b-1\) means are equal but differ from the first mean \(\mu_{1}\), provided that \(d \neq 0\). Let independent random samples of size \(a\) be taken from the \(b\) normal distributions with common unknown variance \(\sigma^{2}\). (a) Show that the maximum likelihood estimators of \(\mu\) and \(d\) are \(\hat{\mu}=\bar{X} . .\) and $$ \hat{d}=\frac{\sum_{j=2}^{b} \bar{X}_{. j} /(b-1)-\bar{X}_{.1}}{b} $$ (b) Using Exercise \(9.1 .3\), find \(Q_{6}\) and \(Q_{7}=c \hat{d}^{2}\) so that, when \(d=0, Q_{7} / \sigma^{2}\) is \(\chi^{2}(1)\) and $$ \sum_{i=1}^{a} \sum_{j=1}^{b}\left(X_{i j}-\bar{X}_{n}\right)^{2}=Q_{3}+Q_{6}+Q_{7} $$ (c) Argue that the three terms in the right-hand member of Part (b), once divided by \(\sigma^{2}\), are independent random variables with chi-square distributions, provided that \(d=0\). (d) The ratio \(Q_{7} /\left(Q_{3}+Q_{6}\right)\) times what constant has an \(F\) -distribution, provided that \(d=0\) ? Note that this \(F\) is really the square of the two-sample \(T\) used to test the equality of the mean of the first distribution and the common mean of the other distributions, in which the last \(b-1\) samples are combined into one.

Let \(Q=X_{1} X_{2}-X_{3} X_{4}\), where \(X_{1}, X_{2}, X_{3}, X_{4}\) is a random sample of size 4 from a distribution which is \(N\left(0, \sigma^{2}\right) .\) Show that \(Q / \sigma^{2}\) does not have a chi-square distribution. Find the mgf of \(Q / \sigma^{2}\).

Using the background of the two-way classification with one observation per cell, show that the maximum likelihood estimator of \(\alpha_{i}, \beta_{j}\), and \(\mu\) are \(\hat{\alpha}_{i}=\bar{X}_{i .}-\bar{X}_{. .}\) \(\hat{\beta}_{j}=\bar{X}_{. j}-\bar{X}_{. .}\), and \(\hat{\mu}=\bar{X}_{. .}\), respectively. Show that these are unbiased estimators of their respective parameters and compute \(\operatorname{var}\left(\hat{\alpha}_{i}\right), \operatorname{var}\left(\hat{\beta}_{j}\right)\), and \(\operatorname{var}(\hat{\mu})\).

Let \(X_{1}, X_{2}, X_{3}, X_{4}\) be a random sample of size \(n=4\) from the normal distribution \(N(0,1) .\) Show that \(\sum_{i=1}^{4}\left(X_{i}-\bar{X}\right)^{2}\) equals $$ \frac{\left(X_{1}-X_{2}\right)^{2}}{2}+\frac{\left[X_{3}-\left(X_{1}+X_{2}\right) / 2\right]^{2}}{3 / 2}+\frac{\left[X_{4}-\left(X_{1}+X_{2}+X_{3}\right) / 3\right]^{2}}{4 / 3} $$ and argue that these three terms are independent, each with a chi-square distribution with 1 degree of freedom.

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