Home Visual Guide to Statistics. Part I: Basics of Point Estimation

Visual Guide to Statistics. Part I: Basics of Point Estimation

This series of posts is a guidance for those who already have knowledge in probability theory and would like to become familiar with mathematical statistics. Basically, these are notes from lectures I attended while being a student in Christian-Albrechts University in Kiel, Germany. They helped me close all the gaps in my knowledge of math under the hood of modern statistics. For those who are interested in the lectures themselves can refer to the original material or my translation to Russian.

This post in particular focuses on point estimators of distribution parameters and their characteristics.


Imagine that you are a pharmaceutical company, which is about to introduce a new drug into production. Prior to launch you need to carry out experiments to assess its quality depending on the dosage. Say you give this medicine to an animal, after which the animal is examined and checked whether it has recovered or not by taking a dose of $X$. You can think of the result as random variable $Y$ following Bernoulli distribution:

\[Y \sim \operatorname{Bin}(1, p(X)),\]

where $p(X)$ is a probability of healing given dose $X$.

Typically, several independent experiments $Y_1, \dots, Y_n$ with different doses $X_1, \dots, X_n$ are made, such that

\[Y_i \sim \operatorname{Bin}(1, p(X_i)).\]

Our goal is to estimate function $p: [0, \infty) \rightarrow [0, 1]$. For example, we can simplify to parametric model

\[p(x) = 1 - e^{-\vartheta x}, \quad \vartheta > 0.\]

Then estimating $p(x)$ is equal to estimating parameter $\vartheta $.

Fig. 1. Visualization of statistical experiments. The question arises: how do we estimate the value of $\vartheta$ based on our observations?

Formally, we can define parameter space $\Theta$ with $\vert \Theta \vert \geq 2$ and family of probability measures $\mathcal{P} = \lbrace P_\vartheta \mid \vartheta \in \Theta \rbrace$, where $P_\vartheta \neq P_{\vartheta’} \ \forall \vartheta \neq \vartheta’$. Then we are interested in the true distribution $P \in \mathcal{P}$ of random variable $X$.

Recall from probability theory that random variable $X$ is a mapping from set of all possible outcomes $\Omega$ to a sample space $\mathcal{X}$. On the basis of given sample $x = X(\omega)$, $\omega \in \Omega$ we make a decision about the unknown $P$. By identifying family $\mathcal{P}$ with the parameter space $\Theta$, a decision for $P$ is equivalent to a decision for $\vartheta$. In our example above

\[Y_i \sim \operatorname{Bin}(1, 1 - e^{-\vartheta X_i}) = P_\vartheta^i\]


\[\mathcal{X} = \{0, 1\}^n, \quad \Theta=\left[0, \infty\right), \quad \mathcal{P}=\{\otimes_{i=1}^nP_{\vartheta}^i \mid \vartheta>0 \}.\]

Uniformly best estimator

Mandatory parameter estimation example which can be found in every statistics handbook is mean and variance estimation for Normal distribution. Let $X_1, \dots, X_n$ i.i.d. $\sim \mathcal{N}(\mu, \sigma^2) = P_{\mu, \sigma^2}$. The typical estimation for $\vartheta = (\mu, \sigma^2)$ would be

\[g(x) = \begin{pmatrix} \overline{x}_n \\ \hat{s}_n^2 \end{pmatrix} = \begin{pmatrix} \frac{1}{n} \sum_{i=1}^n x_i \\ \frac{1}{n} \sum_{i=1}^n (x_i-\overline{x}_n)^2 \end{pmatrix}.\]

We will get back to characteristics of this estimation later. But now it is worth noting that we are not always interested in $\vartheta$ itself, but in an appropriate functional $\gamma(\vartheta)$. We can see it in another example.

Let $X_1, \dots, X_n$ i.i.d. $\sim F$, where $F(x) = \mathbb{P}(X \leq x)$ is unknown distribution function. Here $\Theta$ is an infinite-dimensional family of distribution functions. Say we are interested in value of this function at point $k$:

\[\gamma(F) = F(k).\]

Then a point estimator could be $g(x) = \frac{1}{n} \sum_{i=1}^n \mathbf{1}_{\lbrace X_i \leq k \rbrace }$.

Now we are ready to construct formal definition of parameter estimation. Let’s define measurable space $\Gamma$ and mapping $\gamma: \Theta \rightarrow \Gamma$. Then measurable function $ g: \mathcal{X} \rightarrow \Gamma $ is called (point) estimation of $\gamma(\vartheta)$.

But how do we choose point estimator and how we can measure its goodness? Let’s define a criteria, non-negative function $L: \Gamma \times \Gamma \rightarrow [0, \infty)$, which we will call loss function, and for estimator $g$ function

\[R(\vartheta, g) = \mathbb{E}[L(\gamma(\vartheta), g(X))] = \int_\mathcal{X} L(\gamma(\vartheta), g(X)) P_\vartheta(dx)\]

we will call the risk of $g$ under $L$.

If $\vartheta$ is the true parameter and $g(x)$ is an estimation, then $L(\gamma(\vartheta), g(x))$ measures the corresponding loss. If $\Gamma$ is a metric space, then loss functions typically depend on the distance between $\gamma(\vartheta)$ and $g(x)$, like the quadratic loss $L(x, y)=(x-y)^2$ for $\Gamma = \mathbb{R}$. The risk then is the expected loss.

Suppose we have a set of all possible estimators $g$ called $\mathcal{K}$. Then it is natural to search for an estimator, which mimimizes our risk, namely $\tilde{g} \in \mathcal{K}$, such that

\[R(\vartheta, \tilde{g}) = \inf_{g \in \mathcal{K}} R(\vartheta, g), \quad \forall \vartheta \in \Theta.\]

Let’s call $\tilde{g}$ an uniformly best estimator.

Sadly, in general, neither uniformly best estimators exist nor is one estimator uniformly better than another. For example, let’s take normal random variable with unit variance and estimate its mean $\gamma(\mu) = \mu$ with quadratic loss. Pick the trivial constant estimator $g_\nu(x)=\nu$. Then

\[R(\mu, g_\nu) = \mathbb{E}[(\mu - \nu)^2] = (\mu - \nu)^2.\]

In particular, $R(\nu, g_\nu)=0$. Thus no $g_\nu$ is uniformly better than some $g_\mu$. Also, in order to obtain a uniformly better estimator $\tilde{g}$,

\[\mathbb{E}[(\tilde{g}(X)-\mu)^2]=0 \quad \forall \mu \in \mathbb{R}\]

has to hold, which basically means that $\tilde{g}(x) = \mu$ with probability $1$ for every $\mu \in \mathbb{R}$, which of course is impossible.

UMVU estimator

In order to still get optimal estimators we have to choose other criteria than a uniformly smaller risk. What should be our objective properties of $g$?

Let’s think of difference between this estimator’s expected value and the true value of $\gamma$ being estimated:

\[B_\vartheta(g) = \mathbb{E}[g(X)] - \gamma(\vartheta).\]

This value in is called bias of $g$ and estimator $g$ is called unbiased if

\[B_\vartheta(g) = 0 \quad \forall \vartheta \in \Theta.\]

It is reasonable (at least at the start) to put constraint on unbiasedness for $g$ and search only in

\[\mathcal{E}_\gamma = \lbrace g \in \mathcal{K} \mid B_\vartheta(g) = 0 \rbrace.\]

Surely there can be infinite number of unbiased estimators, and we not only interested in expected value of $g$, but also in how $g$ can vary from it. Variance of $g$ can be chosen as our metric for goodness. We call estimator $\tilde{g}$ uniformly minimum variance unbiased (UMVU) if

\[\operatorname{Var}(\tilde{g}(X)) = \mathbb{E}[(\tilde{g}(X) - \gamma(\theta))^2] = \inf_{g \in \mathcal{E}_\gamma} \operatorname{Var}(g(X)).\]

In general, if we choose $L(x, y) = (x - y)^2$, then

\[MSE_\vartheta(g) = R(\vartheta, g)=\mathbb{E}[(g(X)-\gamma(\vartheta))^2]=\operatorname{Var}_\vartheta(g(X))+B_\vartheta^2(g)\]

is called the mean squared error. Note that in some cases biased estimators have lower MSE because they have a smaller variance than does any unbiased estimator.

Chi-squared and t-distributions

Remember we talked about $\overline{x}_n$ and $\hat{s}_n^2$ being typical estimators for mean and standard deviation of normally distributed random variable? Now we are ready to talk about their properties, but firstly we have to introduce two distributions:

  • Let $X_1, \dots, X_n$ be i.i.d. $\sim \mathcal{N}(0, 1)$. Then random variable $Z = \sum_{i=1}^n X_i^2$ has chi-squared distribution with $n$ degrees of freedom (notation: $Z \sim \chi_n^2$). Its density:

    \[f_{\chi_n^2}(x) = \frac{x^{\frac{n}{2}-1} e^{-\frac{x}{2}}}{2^{\frac{n}{2}}\Gamma\big(\frac{n}{2}\big)}, \quad x > 0,\]

    where $\Gamma(\cdot)$ is a gamma function:

    \[\Gamma(\alpha) = \int_0^\infty x^{\alpha-1} e^{-x} dx, \quad \alpha > 0.\]

    It’s easy to see that $\mathbb{E}[Z] = \sum_{i=1}^n \mathbb{E}[X_i^2] = n$ and

    \[\operatorname{Var}(Z) = \sum_{i=1}^n \operatorname{Var}(X_i^2) = n\big(\mathbb{E}[X_1^4]) - \mathbb{E}[X_1^2]^2\big) = 2n.\]
  • Let $Y \sim \mathcal{N}(0, 1)$ and $Z \sim \chi_n^2$, then

    \[T = \frac{Y}{\sqrt{Z/n}}\]

    has t-distribution with $n$ degrees of freedom (notation $T \sim t_n$). Its density:

    \[f_{t_n}(x) = \frac{\Gamma \big( \frac{n+1}{2} \big) } { \sqrt{n \pi} \Gamma \big( \frac{n}{2} \big) } \Big( 1 + \frac{x^2}{n} \Big)^{\frac{n+1}{2}}.\]

Fig. 2. Probability density functions for $\chi_n^2$ and $t_n$-distributions. Move slider to observe how they look for different degrees of freedom $n$. Note that with large $n$ $t_n$ converges to normal distribution.

It can now be shown that

\[\overline{X}_n = \frac{1}{n} \sum_{i=1}^n X_i \sim \mathcal{N} \Big( \mu, \frac{\sigma^2}{n} \Big)\]


\[\hat{s}_n^2(X) = \frac{1}{n}\sum_{i=1}^n (X_i - \overline{X}_n)^2 \sim \frac{\sigma^2}{n} \chi^2_{n-1}.\]

As a consequence:

\[\frac{(n-1)(\overline{X}_n-\mu)}{\sqrt{n}s_n^2(X)} \sim t_{n-1}.\]
Proof Distribution of $\overline{X}_n$ follows from properties of Normal distribution. Let $$ Y_i = \frac{X_i - \mu}{\sigma} \sim \mathcal{N}(0, 1)$$ and $Y = (Y_1, \dots, Y_n)^T$. Choose orthogonal matrix $A$ such that its last row: $$ v^T = \Big( \frac{1}{\sqrt{n}} \dots \frac{1}{\sqrt{n}} \Big).$$ Then for $Z = AY$ the following equality holds: $$ \sum_{i=1}^n Z_i^2 = Z^TZ = Y^TA^TAY = Y^TY= \sum_{i=1}^n Y_i^2.$$ From $\operatorname{Cov}(Z)=A^TA = \mathbb{I}_n$ we have $Z \sim \mathcal{N}(0, \mathbb{I}_n).$ Also $$ \begin{aligned} \sqrt{n} \overline{X}_n &= \frac{1}{\sqrt{n}} \sum_{i=1}^n (\sigma Y_i + \mu) \\ & = \sigma v^T Y + \sqrt{n} \mu \\ &= \sigma Z_n + \sqrt{n} \mu \end{aligned} $$ and $$ \begin{aligned} n \hat{s}_n^2(X) &= \sum_{i=1}^n (X_i - \overline{X}_n)^2 = \sigma^2 \sum_{i=1}^n(Y_i - \overline{Y}_n)^2 \\ & = \sigma^2 \big(\sum_{i=1}^n Y_i^2 - n \overline{Y}_n^2\big) = \sigma^2 \big(\sum_{i=1}^n Y_i^2 - \big(\frac{1}{n} \sum_{i=1}^n Y_i^2 \big)^2 \big) \\ & = \sigma^2 (\sum_{i=1}^n Z_i^2 - Z_n^2) = \sigma^2 \sum_{i=1}^{n-1} Z_i^2 \sim \chi_{n-1}^2. \end{aligned} $$ Both estimators are independent as functions of $Z_n$ and $Z_1, \dots, Z_{n-1}$ respectively.

Let’s check which of these estimators are unbiased. We have $\mathbb{E}[\overline{X}_n] = \mu$, therefore $\overline{X}_n$ is unbiased. On the other hand

\[\mathbb{E}[\hat{s}_n^2(X)] = \frac{\sigma^2}{n} (n - 1) \neq \sigma^2.\]

Fig. 3. Statistical experiments in estimating $\sigma$ for $X_1, \dots, X_n$ i.i.d. $\sim \mathcal{N}(0, 1)$. We see here that while $\overline{X}_n$ varies around $\mu=0$, expected value of estimator $\hat{s}_n^2(X)$ is lower than $\sigma^2 = 1$.

So far we figured the unbiasedness of $g(X) = \overline{X}_n$. But how can we tell if $\overline{X}_n$ is an UMVU estimator? Can we find an estimator of $\mu$ with variance lower than $\frac{\sigma^2}{n}$?

Efficient estimator

Given a set of unbiased estimators, it is not an easy task to determine which one provides the smallest variance. Luckily, we have a theorem which gives us a lower bound for an estimator variance.

Suppose we have a family of densities $f(\cdot, \vartheta)$, such that following regularity conditions are satisfied:

  • Set $M_f=\lbrace x \in \mathcal{X} \mid f(x, \vartheta) > 0 \rbrace$ doesn’t depend on $\vartheta$
  • Partial derivative $\frac{\partial}{\partial \vartheta} \log f(x, \vartheta)$ exists $\forall x \in \mathcal{X}$.
  • The following equalities hold: 1
    • $\mathbb{E} \big[\frac{\partial}{\partial \vartheta} \log f(X, \vartheta)\big] = 0,$
    • $\mathbb{E} \big[g(X) \frac{\partial}{\partial \vartheta} \log f(X, \vartheta)\big] = \frac{\partial}{\partial \vartheta} \mathbb{E}[g(X)].$
  • $0<\mathbb{E} \big[\big(\frac{\partial}{\partial \vartheta} \log f(X, \vartheta)\big)^2\big]<\infty$

Let’s define functions

\[U_\vartheta(x) = \left\{\begin{array}{ll} \frac{\partial}{\partial \vartheta} \log f(x, \vartheta), & \text{if } x \in M_f, \\ 0, & \text{otherwise,} \end{array} \right.\]


\[\mathcal{I}(f(\cdot, \vartheta))=\mathbb{E} \big[\big(\frac{\partial}{\partial \vartheta} \log f(X, \vartheta)\big)^2\big].\]

Under given regularity conditions we have

\[\mathbb{E}[U_\vartheta(X)] = \mathbb{E}\big[\frac{\partial}{\partial \vartheta} \log f(x, \vartheta)\big] = \frac{\partial}{\partial \vartheta} \mathbb{E}[\log f(x, \vartheta)] = 0\]


\[\operatorname{Var}(U_\vartheta(X)) = \mathbb{E}[(U_\vartheta(X))^2]=\mathcal{I}(f(\cdot, \vartheta)).\]

Then using Cauchy-Schwartz inequality we get

\[\begin{aligned} \big( \frac{\partial}{\partial \vartheta} \mathbb{E}[g(X)] \big)^2 &= \big( \mathbb{E}[g(X) \cdot U_\vartheta(X)] \big)^2 \\ & = \big(\operatorname{Cov}(g(X), U_\vartheta(X)) \big)^2 \\ & \leq \operatorname{Var}(g(X))\cdot \operatorname{Var}(U_\vartheta(X)) \\ &= \mathcal{I}(f(\cdot, \vartheta))\cdot \operatorname{Var}(g(X)). \end{aligned}\]

The resulting inequality:

\[\operatorname{Var}(g(X)) \geq \frac{\big(\frac{\partial}{\partial \vartheta} \mathbb{E}[g(X)]\big)^2}{\mathcal{I}(f(\cdot, \vartheta))} \quad \forall \vartheta \in \Theta\]

gives us Cramér–Rao bound. Function $\mathcal{I}(f(\cdot, \vartheta))$ is called Fisher information for family $\mathcal{P} = \lbrace P_\vartheta \mid \vartheta \in \Theta \rbrace$. If an unbiased estimator $g$ satisfies the upper equation with equality, then it is called efficient.

This theorem gives a lower bound for the variance of an estimator for $\gamma(\vartheta) = \mathbb{E}[g(X)]$ and can be used in principle to obtain UMVU estimators. Whenever the regularity conditions are satisfied for all $g \in \mathcal{E}_\gamma$, then any efficient and unbiased estimator is UMVU.

Also, for a set of i.i.d. variables $X_1, \dots X_n$, meaning that their joint density distribution is

\[f(x,\vartheta) = \prod_{i=1}^n f^i(x,\vartheta),\]

we have

\[\mathcal{I}(f(\cdot, \vartheta))=n\mathcal{I}(f^1(\cdot, \vartheta)).\]

Let’s get back to the example with $X_1, \dots, X_n$ i.i.d. $\sim \mathcal{N}(\mu, 1)$ having the density

\[f^1(x, \vartheta) = \frac{1}{\sqrt{2\pi}} e^{-\frac{(x-\mu)^2}{2}}.\]


\[\mathcal{I}(f^1(\cdot, \mu)) = \mathbb{E} \Big[ \big( \frac{\partial}{\partial \mu} \log f^1 (X_1, \mu)\big)^2 \Big] = \mathbb{E}[(X_1 - \mu)^2] = 1.\]

In particular, for $X = (X_1, \dots, X_n)$ Fisher information $\mathcal{I}(f(X, \mu)) = n$ and Cramér–Rao bound for unbiased estimator:

\[\operatorname{Var}(g(X)) \geq \frac{1}{n} \big( \frac{\partial}{\partial \mu} \mathbb{E}[g(X)] \big)^2 = \frac{1}{n}.\]

Therefore, $g(x) = \overline{x}_n$ is an UMVU estimator.

Multidimensional Cramér–Rao inequality

Define function

\[G(\vartheta)=\Big( \frac{\partial}{\partial \vartheta_j} \mathbb{E}[g_i(X)] \Big)_{i,j} \in \mathbb{R}^{k \times d}.\]

Then with multidimensional Cauchy-Shwartz inequality one can prove that under similar regularity conditions we have:

\[\operatorname{Cov}(g(X)) \geq G(\vartheta) \mathcal{I}^{-1}(f(\cdot, \vartheta))G^T(\vartheta) \in \mathbb{R}^{k \times k},\]

in the sense of Löwner ordering2, where

\[\mathcal{I}(f(\cdot, \vartheta))=\Big( \mathbb{E}\Big[\frac{\partial}{\partial \vartheta_i} \log f(X, \vartheta) \cdot \frac{\partial}{\partial \vartheta_j} \log f(X, \vartheta) \Big] \Big)_{i,j=1}^d \in \mathbb{R}^{d \times d}.\]

For an example with $X_1, \dots X_n$ i.i.d. $\sim \mathcal{N}(\mu, \sigma^2)$ with density

\[f^1(x,\vartheta)=\frac{1}{\sqrt{2\pi \sigma^2}} \exp \Big(-\frac{(x-\mu)^2}{2\sigma^2}\Big)\]

we have

\[U_\vartheta = \Big(\frac{\partial}{\partial \mu} \log f^1(X_1,\vartheta), \frac{\partial}{\partial \sigma^2} \log f^1(X_1,\vartheta)\Big)^T = \begin{pmatrix} (X_1-\mu)/\sigma^2 \\ -\frac{1}{2\sigma^2}+\frac{1}{\sigma^4}(X_1-\mu)^2 \end{pmatrix}.\]

Fisher information then

\[\mathcal{I}(f^1(\cdot, \vartheta))=\mathbb{E}[U_\vartheta U_\vartheta^T]= \begin{pmatrix} \sigma^{-2} & 0 \\ 0 & \frac{1}{2}\sigma^{-4} \end{pmatrix} = \frac{1}{n}\mathcal{I}(f(\cdot, \vartheta)).\]

If $g(X)$ is an unbiased estimator, then $G(\vartheta)$ is identity matrix and Cramér–Rao bound then

\[\begin{aligned} \operatorname{Cov}_\vartheta(g(X)) & \geq G(\vartheta) \ \mathcal{I}^{-1} (f(\cdot, \vartheta)) \ G^T(\vartheta) \\ &= \mathcal{I}^{-1}(f(\cdot, \vartheta)) = \begin{pmatrix} \frac{\sigma^{2}}{n} & 0 \\ 0 & \frac{2\sigma^{4}}{n} \end{pmatrix}. \end{aligned}\]

In particular for an unbiased estimator

\[\widetilde{g}(X)=\Big(\overline{X}_n, \frac{1}{n-1} \sum_{i=1}^n(X_j-\overline{X}_n)^2 \Big)^T\]

the following inequality holds

\[\operatorname{Cov}_\vartheta(\widetilde{g}(X)) = \begin{pmatrix} \frac{\sigma^{2}}{n} & 0 \\ 0 & \frac{2\sigma^{4}}{n-1} \end{pmatrix} \geq \mathcal{I}(f(\cdot, \vartheta)),\]

therefore $\widetilde{g}$ is not efficient.

Exponential family

In the previous examples, we consider without proof the fulfillment of all regularity conditions of the Cramér–Rao inequality. Next, we will discuss a family of distributions for which the Cramér–Rao inequality turns into an equality.

Proposition: let $P_\vartheta$ be distribution with density

\[f(x, \vartheta) = c(\vartheta) h(x) \exp(\vartheta T(x)) \quad \forall \vartheta \in \Theta.\]

Then equality in Cramér–Rao theorem holds for $g(x) = T(x)$.

Proof First let us note that $\int_{\mathcal{X}}f(x)\mu(dx) = 1$ for all $\vartheta \in \Theta$, hence $$ c(\vartheta)=\Big( \int_{\mathcal{X}} h(x)\exp (\vartheta T(x) ) dx \Big)^{-1} $$ and $$ \begin{aligned} 0 & = \frac{\partial}{\partial \vartheta} \int_{\mathcal{X}} c(\vartheta) h(x) \exp ( \vartheta T(x) ) dx \\ & = \int_{\mathcal{X}} (c'(\vartheta)+c(\vartheta)T(x)) h(x) \exp ( \vartheta T(x) ) dx. \end{aligned} $$ Using these two equations we get $$ \begin{aligned} \mathbb{E}[T(X)] & = c(\vartheta) \int_{\mathcal{X}} h(x) T(x) \exp ( \vartheta T(x)) dx \\ & = -c'(\vartheta) \int_{\mathcal{X}}h(x) \exp ( \vartheta T(x) ) dx \\ & = -\frac{c'(\vartheta)}{c(\vartheta)}=(-\log c(\vartheta))'. \end{aligned} $$ Fisher information: $$ \mathcal{I}(f(\cdot, \vartheta)) = \mathbb{E}\Big[\Big( \frac{\partial}{\partial \vartheta} \log f(X, \vartheta) \Big)^2\Big]=\mathbb{E}[(T(X)+(\log c(\vartheta))')^2]=\operatorname{Var}(T(X)). $$ Also $$ \begin{aligned} \frac{\partial}{\partial \vartheta} \mathbb{E}[T(X)] & =\int_{\mathcal{X}} c'(\vartheta) h(x) T(x) \exp ( \vartheta T(x) ) dx + \int_{\mathcal{X}} c(\vartheta) h(x) T^2(x) \exp ( \vartheta T(x) ) dx \\ & = \frac{c'(\vartheta)}{c(\vartheta)} \int_{\mathcal{X}} c(\vartheta) h(x) T(x) \exp ( \vartheta T(x) ) dx + \mathbb{E}[(T(X))^2] \\ & = \mathbb{E}[(T(X))^2] - (\mathbb{E}[T(X)])^2. \end{aligned} $$ Therefore, $$ \frac{\Big(\frac{\partial}{\partial\vartheta}\mathbb{E}[T(X)] \Big)^2}{\mathcal{I}(f(\cdot, \vartheta))}= \operatorname{Var}(T(X)). $$

Formally, family $\mathcal{P} = \lbrace P_\vartheta \mid \vartheta \in \Theta \rbrace $ is called an exponential family if there exist mappings $c, Q_1, \dots Q_k: \Theta \rightarrow \mathbb{R}$ and $h, T_1, \dots T_k: \mathcal{X} \rightarrow \mathbb{R}$ such that

\[f(x,\vartheta) = c(\vartheta) h(x) \exp \Big( \sum_{j=1}^k Q_j(\vartheta) T_j(x) \Big).\]

$\mathcal{P}$ is called $k$-parametric exponential family if functions $1, Q_1, \dots Q_k$ and $1, T_1, \dots T_k$ are linearly independent. Then we have equality to Cramér–Rao bound for $g = (T_1, \dots T_k)^T$.

Here are some examples:

  • If $X \sim \operatorname{Bin}(n, \vartheta)$, then

    \[\begin{aligned} f(x, \vartheta) &= \binom n x \vartheta^x (1-\vartheta)^{n-x} \\ &= (1-\vartheta)^n \binom n x \exp \Big(x \log \frac{\vartheta}{1-\vartheta} \Big). \end{aligned}\]

    Here $c(\vartheta) = (1-\vartheta)^n$, $h(x) = \binom n x$, $T_1(x) = x$ and $Q_1(\vartheta) = \log \frac{\vartheta}{1-\vartheta}$.

  • If $X \sim \mathcal{N}(\mu, \sigma^2)$, then $\vartheta = (\mu, \sigma^2)^T$ and

    \[\begin{aligned} f(x, \vartheta) &= \frac{1}{\sqrt{2\pi\sigma^2}} \exp\Big( \frac{(x-\mu)^2}{2\sigma^2} \Big) \\ &= \frac{1}{\sqrt{2\pi\sigma^2}} \exp \Big( -\frac{\mu^2}{2\sigma^2} \Big) \exp\Big( -\frac{x^2}{2\sigma^2} + \frac{\mu x}{\sigma^2} \Big), \end{aligned}\]

    where $c(\vartheta) = \frac{1}{\sqrt{2\pi\sigma^2}} \exp \big( -\frac{\mu^2}{2\sigma^2} \big) $, $Q_1(\vartheta) = -\frac{1}{2\sigma^2}$, $Q_2(\vartheta) = \frac{\mu}{\sigma^2}$, $T_1(x)=x^2$ and $T_2(x)=x$.

  • If $X \sim \operatorname{Poisson}(\lambda)$, then

\[f(x, \lambda) = \frac{\lambda^x e^{-\lambda}}{x!} = e^{-\lambda} \frac{1}{x!} \exp \big(x \log \lambda \big).\]

Denoting $Q(\vartheta) = (Q_1(\vartheta), \dots, Q_k(\vartheta))^T$ we get transformed parametric space $ \Theta^* = Q(\Theta) $, which we call natural parametric space. In examples above

  • $X \sim \operatorname{Bin}(n, \vartheta)$: $\Theta^* = \lbrace \log \frac{\vartheta}{1-\vartheta} \mid \vartheta \in (0, 1) \rbrace = \mathbb{R}$.
  • $X \sim \mathcal{N}(\mu, \sigma^2)$: $\Theta^* = \big\lbrace \big( \frac{\mu}{\sigma^2}, -\frac{1}{\sigma^2} \big) \mid \mu \in \mathbb{R}, \sigma^2 \in \mathbb{R}^+ \big\rbrace = \mathbb{R} \times \mathbb{R}^-.$
  • $X \sim \operatorname{Poisson}(\lambda)$: $\Theta^* = \lbrace \log \lambda \mid \lambda \in \mathbb{R}^+ \rbrace = \mathbb{R}$.

It must be noted that for an exponential family $\mathcal{P}$ estimator $T(X) = (T_1(X), \dots T_k(X))$ is UMVU for $\mathbb{E}[T(X)]$. For example, if $X_1, \dots X_n$ i.i.d. $\sim \mathcal{N}(\mu, \sigma^2)$ with joint density

\[f(x,\vartheta) = c(\vartheta) \exp \Big( -\frac{n}{2\sigma^2}\Big( \frac{1}{n} \sum_{i=1}^n x_i^2 \Big) + \frac{n\mu}{\sigma^2}\Big( \frac{1}{n}x_i \Big) \Big),\]

then estimator

\[T(X) = \Big( \frac{1}{n} \sum_{i=1}^n X_i, \frac{1}{n} \sum_{i=1}^n X_i^2 \Big)\]

is efficient for $(\mu, \mu^2 + \sigma^2)^T$.

Common estimation methods

If distribution doesn’t belong to exponential family, then for such case there exist two classical estimation methods:

  • Method of moments. Let $X_1, \dots X_n$ i.i.d. $\sim P_\vartheta$ and

    \[\gamma(\vartheta) = f(m_1, \dots, m_k),\]

    where $m_j = \mathbb{E}[X_1^j]$. Then estimation by method of moments will be

    \[\hat{\gamma} (X) = f(\hat{m}_1, \dots, \hat{m}_k),\]

    where $m_j = \frac{1}{n}\sum_{i=1}^nX_i^j$.

  • Maximum likelihood method. Say $\gamma(\vartheta) = \vartheta \in \mathbb{R}^k$. Then $\hat{\vartheta}(x)$ is a maximum likelihood estimator if

    \[f(x, \hat{\vartheta}) = \sup_{\vartheta \in \Theta} f(x, \vartheta).\]

Again in example $X_1, \dots X_n$ i.i.d. $\sim \mathcal {N}(\mu, \sigma^2)$ an estimator for $\vartheta = (\mu, \sigma^2)^T = (m_1, m_2 - m_1^2)^T$ by method of moments will be

\[\hat{\gamma}(\vartheta)=(\hat{m}_1, \hat{m}_2-\hat{m}_1^2)^T=(\overline{x}_n, \hat{s}_n^2)^T.\]

It’s easy to show that this estimator coincides with the estimation obtained by the maximum likelihood method.

Let’s take another example, $X_1, \dots X_n$ i.i.d. $\sim \mathcal{U}(0, \vartheta)$, where estimated parameter $\vartheta > 0$. One can show that estimator

\[g_{ML}(X) = X_{(n)} = \max \lbrace X_1, \dots X_n \rbrace\]

is a maximum-likelihood estimator. On the other hand,

\[g_{MM}(X) = 2 \overline{X}_n\]

is an estimator by method of moments. Also, maximum-likelihood estimator follows scaled Beta-distribution, $g_{ML}(X) \sim \vartheta B(n, 1)$, and therefore it is biased:

\[\mathbb{E}[g_{ML}(X)] = \vartheta\frac{n}{n+1}.\]

UMVU estimator is $g(X) = X_{(n)} (1 + \frac{1}{n})$, and its variance:

\[\operatorname{Var}[g(X)] = \vartheta^2\frac{1}{n(n+2)} < \frac{\vartheta^2}{n}\]

However, the Cramér-Rao lower bound is $\frac{\vartheta^2}{n}$. This shows importance of regularity conditions for Cramér-Rao theorem. Here, invariance of $M_f$ is not satisified and Cramér-Rao inequality doesn’t hold.

  1. Let’s rewrite these equations in equivalent forms:

    \[\int_\mathcal{X} \frac{\partial}{\partial \vartheta} \log f(x, \vartheta) f(x, \vartheta) d x = \int_\mathcal{X} \frac{\partial}{\partial \vartheta} f(x, \vartheta) d x=\frac{\partial}{\partial \vartheta}\int_\mathcal{X} f(x, \vartheta) d x =0,\] \[\int_\mathcal{X} g(x) \frac{\partial}{\partial \vartheta} f(x, \vartheta) dx = \frac{\partial}{\partial \vartheta} \int_\mathcal{X} g(x) f(x,\vartheta) dx.\]

    In both cases it means that we can interchange differentiation and integration. 

  2. For symmetric matrices $A$ and $B$ we say that

    \[A \geq 0 \Longleftrightarrow A \text{ is positive semi-definite},\] \[A \geq B \Longleftrightarrow A - B \geq 0.\]

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