# Lehmann–Scheffé theorem

In statistics, the Lehmann–Scheffé theorem is a prominent statement, tying together the ideas of completeness, sufficiency, uniqueness, and best unbiased estimation.[1] The theorem states that any estimator which is unbiased for a given unknown quantity and that depends on the data only through a complete, sufficient statistic is the unique best unbiased estimator of that quantity. The Lehmann–Scheffé theorem is named after Erich Leo Lehmann and Henry Scheffé, given their two early papers.[2][3]

If T is a complete sufficient statistic for θ and E(g(T)) = τ(θ) then g(T) is the uniformly minimum-variance unbiased estimator (UMVUE) of τ(θ).

## Statement

Let ${\displaystyle {\vec {X}}=X_{1},X_{2},\dots ,X_{n}}$ be a random sample from a distribution that has p.d.f (or p.m.f in the discrete case) ${\displaystyle f(x:\theta )}$ where ${\displaystyle \theta \in \Omega }$ is a parameter in the parameter space. Suppose ${\displaystyle Y=u({\vec {X}})}$ is a sufficient statistic for θ, and let ${\displaystyle \{f_{Y}(y:\theta ):\theta \in \Omega \}}$ be a complete family. If ${\displaystyle \varphi :\operatorname {E} [\varphi (Y)]=\theta }$ then ${\displaystyle \varphi (Y)}$ is the unique MVUE of θ.

### Proof

By the Rao–Blackwell theorem, if ${\displaystyle Z}$ is an unbiased estimator of θ then ${\displaystyle \varphi (Y):=\operatorname {E} [Z\mid Y]}$ defines an unbiased estimator of θ with the property that its variance is not greater than that of ${\displaystyle Z}$.

Now we show that this function is unique. Suppose ${\displaystyle W}$ is another candidate MVUE estimator of θ. Then again ${\displaystyle \psi (Y):=\operatorname {E} [W\mid Y]}$ defines an unbiased estimator of θ with the property that its variance is not greater than that of ${\displaystyle W}$. Then

${\displaystyle \operatorname {E} [\varphi (Y)-\psi (Y)]=0,\theta \in \Omega .}$

Since ${\displaystyle \{f_{Y}(y:\theta ):\theta \in \Omega \}}$ is a complete family

${\displaystyle \operatorname {E} [\varphi (Y)-\psi (Y)]=0\implies \varphi (y)-\psi (y)=0,\theta \in \Omega }$

and therefore the function ${\displaystyle \varphi }$ is the unique function of Y with variance not greater than that of any other unbiased estimator. We conclude that ${\displaystyle \varphi (Y)}$ is the MVUE.

## Example for when using a non-complete minimal sufficient statistic

An example of an improvable Rao–Blackwell improvement, when using a minimal sufficient statistic that is not complete, was provided by Galili and Meilijson in 2016.[4] Let ${\displaystyle X_{1},\ldots ,X_{n}}$ be a random sample from a scale-uniform distribution ${\displaystyle X\sim U((1-k)\theta ,(1+k)\theta ),}$ with unknown mean ${\displaystyle \operatorname {E} [X]=\theta }$ and known design parameter ${\displaystyle k\in (0,1)}$. In the search for "best" possible unbiased estimators for ${\displaystyle \theta }$, it is natural to consider ${\displaystyle X_{1}}$ as an initial (crude) unbiased estimator for ${\displaystyle \theta }$ and then try to improve it. Since ${\displaystyle X_{1}}$ is not a function of ${\displaystyle T=\left(X_{(1)},X_{(n)}\right)}$, the minimal sufficient statistic for ${\displaystyle \theta }$ (where ${\displaystyle X_{(1)}=\min _{i}X_{i}}$ and ${\displaystyle X_{(n)}=\max _{i}X_{i}}$), it may be improved using the Rao–Blackwell theorem as follows:

${\displaystyle {\hat {\theta }}_{RB}=\operatorname {E} _{\theta }[X_{1}\mid X_{(1)},X_{(n)}]={\frac {X_{(1)}+X_{(n)}}{2}}.}$

However, the following unbiased estimator can be shown to have lower variance:

${\displaystyle {\hat {\theta }}_{LV}={\frac {1}{k^{2}{\frac {n-1}{n+1}}+1}}\cdot {\frac {(1-k)X_{(1)}+(1+k)X_{(n)}}{2}}.}$

And in fact, it could be even further improved when using the following estimator:

${\displaystyle {\hat {\theta }}_{\text{BAYES}}={\frac {n+1}{n}}\left[1-{\frac {{\frac {X_{(1)}(1+k)}{X_{(n)}(1-k)}}-1}{\left({\frac {X_{(1)}(1+k)}{X_{(n)}(1-k)}}\right)^{n+1}-1}}\right]{\frac {X_{(n)}}{1+k}}}$