# 31 Likelihood Ratio Tests

## 31.1 General Set-up

Most hypothesis testing procedures can be formulated so that a test statistic, $$S(\boldsymbol{x})$$ is applied to the data $$\boldsymbol{x} = (x_1, x_2, \ldots, x_n)^T$$ so that:

1. $$S(\boldsymbol{x}) \geq 0$$
2. The larger $$S(\boldsymbol{x})$$ is, the more significant the test is (i.e., the more evidence against the null in favor of the alternative)
3. The p-value is $$p(\boldsymbol{x}) = \Pr(S(\boldsymbol{X^*}) \geq S(\boldsymbol{x}))$$ where $$S(\boldsymbol{X^*})$$ is distributed according to the null distribution

## 31.2 Significance Regions

A level $$\alpha$$ test is a signficance rule (i.e., a rule for calling a test statistically significant) that results in a false positive rate (i.e., Type I error rate) of $$\alpha$$. Under our set-up, significance regions take the form:

$\Gamma_\alpha = \left\{\boldsymbol{x}: S(\boldsymbol{x}) \geq c_{1-\alpha} \right\},$

where $$c_{1-\alpha}$$ is the $$(1-\alpha)$$ percentile of $$S(\boldsymbol{X}^*)$$ so that $$\Pr(S(\boldsymbol{X}^*) \geq c_{1-\alpha}) = \alpha$$. We restrict $$0 < \alpha < 1$$.

Note that if $$\alpha' \leq \alpha$$ then $$\Gamma_{\alpha'} \subseteq \Gamma_\alpha$$.

## 31.3 P-values

A p-value can be defined in terms of significance regions:

$p(\boldsymbol{x}) = \min \left\{\alpha: \boldsymbol{x} \in \Gamma_\alpha \right\}$

## 31.4 Example: Wald Test

Consider the hypothesis test, $$H_0: \theta=\theta_0$$ vs $$H_1: \theta \not= \theta_0$$. Let $$\hat{\theta}_n(\boldsymbol{x})$$ be the MLE of $$\theta$$. We have

$S(\boldsymbol{x}) = \frac{\left| \hat{\theta}_n(\boldsymbol{x}) - \theta_0 \right|}{\hat{{\operatorname{se}}}\left(\hat{\theta}_n(\boldsymbol{x})\right)},$

$\Gamma_\alpha = \left\{\boldsymbol{x}: S(\boldsymbol{x}) \geq |z_{\alpha/2}| \right\},$

where $$z_{\alpha/2}$$ is the $$\alpha/2$$ percentile of the Normal$$(0,1)$$ distribution.

## 31.5 Neyman-Pearson Lemma

Suppose we are testing $$H_0: \theta = \theta_0$$ vs $$H_1: \theta = \theta_1$$ where in practice $$\theta_0$$ and $$\theta_1$$ are known, fixed quantities. The most powerful test has statistic and significance regions:

$S(\boldsymbol{x}) = \frac{f(\boldsymbol{x}; \theta_1)}{f(\boldsymbol{x}; \theta_0)} = \frac{L(\theta_1; \boldsymbol{x})}{L(\theta_0; \boldsymbol{x})}$

$\Gamma_\alpha = \left\{\boldsymbol{x}: S(\boldsymbol{x}) \geq c_{1-\alpha} \right\},$

where $$c_{1-\alpha}$$ is the $$(1-\alpha)$$ percentile of $$S(\boldsymbol{X}^*)$$ so that $$\operatorname{Pr}_{\theta_0}(S(\boldsymbol{X}^*) \geq c_{1-\alpha}) = \alpha$$.

## 31.6 Simple vs. Composite Hypotheses

A simple hypothesis is defined in terms of a single value, e.g.,

• $$H_0: \mu=0$$
• $$H_0: p = p_0$$ where $$p_0$$ is a placehold for a known, fixed number in practice
• $$H_1: \lambda=5$$

A composite hypothesis is defined by multiple values, e.g.,

• $$H_0: \mu \leq 0$$ vs $$H_1: \mu > 0$$
• $$H_0: p_1 = p_2$$, where $$p_1$$ and $$p_2$$ are two unknown parameters corresponding to two populations
• $$H_1: \mu \not= 0$$

## 31.7 General Hypothesis Tests

Let $$X_1, X_2, \ldots, X_n \stackrel{{\rm iid}}{\sim} F_\theta$$ where $$\theta \in \Theta$$. Let $$\Theta_0, \Theta_1 \subseteq \Theta$$ so that $$\Theta_0 \cap \Theta_1 = \varnothing$$ and $$\Theta_0 \cup \Theta_1 = \Theta$$. The hypothesis test is:

$$H_0: \theta \in \Theta_0$$ vs $$H_1: \theta \in \Theta_1$$

If $$\Theta_0$$ or $$\Theta_1$$ contain more than one value then the corresponding hypothesis is composite.

## 31.8 Composite $$H_0$$

The significance regions indexed by their level $$\alpha$$ are determined so that:

$\Gamma_\alpha = \left\{\boldsymbol{x}: S(\boldsymbol{x}) \geq c_{1-\alpha} \right\},$

where $$c_{1-\alpha}$$ is such that $\max_{\theta \in \Theta_0} \Pr(S(\boldsymbol{X^*}) \geq c_{1-\alpha}) = \alpha.$

In this case,

\begin{align*} p(\boldsymbol{x}) & = \min \left\{\alpha: \boldsymbol{x} \in \Gamma_\alpha \right\} \\ & = \max_{\theta \in \Theta_0} \operatorname{Pr}_\theta (S(\boldsymbol{X}^*) \geq S(\boldsymbol{x})) \end{align*}

## 31.9 Generalized LRT

Let $$X_1, X_2, \ldots, X_n \stackrel{{\rm iid}}{\sim} F_\theta$$ where $$\theta \in \Theta$$ and we are testing $$H_0: \theta \in \Theta_0$$ vs $$H_1: \theta \in \Theta_1$$.

The generalized LRT utilizes test statistic and significance regions:

$\lambda(\boldsymbol{x}) = \frac{\max_{\theta \in \Theta} L(\theta; \boldsymbol{x})}{\max_{\theta \in \Theta_0} L(\theta; \boldsymbol{x})} = \frac{L\left(\hat{\theta}; \boldsymbol{x}\right)}{L\left(\hat{\theta}_0; \boldsymbol{x}\right)}$

$\Gamma_\alpha = \left\{\boldsymbol{x}: \lambda(\boldsymbol{x}) \geq c_{1-\alpha} \right\}$

## 31.10 Null Distribution of Gen. LRT

The null distribution of $$\lambda(\boldsymbol{x})$$ under “certain regularity assumptions” can be shown to be such that, as $$n \rightarrow \infty$$,

$2 \log \lambda(\boldsymbol{x}) \stackrel{D}{\longrightarrow} \chi^2_v$

where $$v = \operatorname{dim}(\Theta) - \operatorname{dim}(\Theta_0)$$.

The significance regions can be more easily written as $$\Gamma_\alpha = \left\{\boldsymbol{x}: 2 \log \lambda(\boldsymbol{x}) \geq c_{1-\alpha} \right\}$$ where $$c_{1-\alpha}$$ is the $$1-\alpha$$ percentile of the $$\chi^2_v$$ distribution.

## 31.11 Example: Poisson

Let $$X_1, X_2, \ldots, X_n \stackrel{{\rm iid}}{\sim} \mbox{Poisson}(\theta)$$ where $$\theta > 0$$ and we are testing $$H_0: \theta = \theta_0$$ vs $$H_1: \theta \not= \theta_0$$. The unconstrained MLE is $$\hat{\theta} = \overline{x}$$. The generalized LRT statistic

$2 \log \lambda(\boldsymbol{x}) = 2 \log \frac{e^{-n \hat{\theta}} \hat{\theta}^{\sum x_i} }{e^{-n \theta_0} \theta_0^{\sum x_i} } = 2 n \left[ (\theta_0 - \hat{\theta}) - \hat{\theta} \log(\theta_0 / \hat{\theta}) \right]$

which has an asymptotic $$\chi^2_1$$ null distribution.

## 31.12 Example: Normal

Let $$X_1, X_2, \ldots, X_n \stackrel{{\rm iid}}{\sim} \mbox{Normal}(\mu, \sigma^2)$$ and we are testing $$H_0: \mu = \mu_0$$ vs $$H_1: \mu \not= \mu_0$$. The generalized LRT can be applied for multidimensional parameter spaces $$\boldsymbol{\Theta}$$ as well. The statistic, which has asymptotic null distribution $$\chi^2_1$$, is

$2 \log \lambda(\boldsymbol{x}) = 2 \log \left( \frac{\hat{\sigma}^2_0}{\hat{\sigma}^2} \right)^{n/2}$

where

$\hat{\sigma}^2_0 = \frac{\sum_{i=1}^n (x_i - \mu_0)^2}{n}, \quad \hat{\sigma}^2 = \frac{\sum_{i=1}^n (x_i - \overline{x})^2}{n}.$