View source for Statistical estimation

<!--
s0873501.png
$#A+1 = 182 n = 0
$#C+1 = 182 : ~/encyclopedia/old_files/data/S087/S.0807350 Statistical estimation
Automatically converted into TeX, above some diagnostics.
Please remove this comment and the {{TEX|auto}} line below,
if TeX found to be correct.
-->

{{TEX|auto}}
{{TEX|done}}

One of the fundamental parts of mathematical statistics, dedicated to the estimation using random observations of various characteristics of their distribution.

===Example 1.===
Let  $  X _ {1} \dots X _ {n} $
be independent random variables (or observations) with a common unknown distribution  $  {\mathcal P} $
on the straight line. The empirical (sample) distribution  $  {\mathcal P} _ {n}  ^  \star  $
which ascribes the weight  $  1/n $
to every random point  $  X _ {n} $
is a [[Statistical estimator|statistical estimator]] for  $  {\mathcal P} $. 
The empirical moments

$$ 
a _  \nu   =  \int\limits x  ^  \nu   d {\mathcal P} _ {n}  ^  \star   = \ 

\frac{1}{n}
 \sum _ { i= } 1 ^ { n }  X _ {i}  $$

serve as estimators for the moments  $  \alpha _  \nu  = \int x  ^  \nu   d {\mathcal P} $. 
In particular,

$$ 
\overline{X}\;  =  
\frac{1}{n}
 \sum _ { i= } 1 ^ { n }  X _ {i}  $$

is an estimator for the mean, and

$$ 
s  ^ {2}  =  
\frac{1}{n}
 \sum _ { i= } 1 ^ { n }  ( X _ {i} - \overline{X}\; )  ^ {2}
$$

is an estimator for the variance.

==Basic concepts.==
In the general theory of estimation, an observation of  $  X $
is a [[Random element|random element]] with values in a measurable space  $  ( \mathfrak X , \mathfrak A) $, 
whose unknown distribution belongs to a given family of distributions  $  P $. 
The family of distributions can always be parametrized and written in the form  $  \{ { {\mathcal P} _  \theta  } : {\theta \in \Theta } \} $. 
Here the form of dependence on the parameter and the set  $  \Theta $
are assumed to be known. The problem of estimation using an observation  $  X $
of an unknown parameter  $  \theta $
or of the value  $  g( \theta ) $
of a function  $  g $
at the point  $  \theta $
consists of constructing a function  $  \theta  ^  \star  ( X) $
from the observations made, which gives a sufficiently good approximation of  $  \theta $
$  ( g( \theta )) $.

A comparison of estimators is carried out in the following way. Let a non-negative loss function  $  w( y _ {1} ;  y _ {2} ) $
be defined on  $  \Theta \times \Theta $
$  ( g( \Theta ) \times g( \Theta )) $, 
the sense of this being that the use of  $  \theta  ^  \star  $
for the actual value of  $  \theta $
leads to losses  $  w( \theta  ^  \star  ;  \theta ) $. 
The mean losses and the risk function  $  R _ {w} ( \theta  ^  \star  ;  \theta ) = {\mathsf E} _  \theta  w( \theta  ^  \star  ;  \theta ) $
are taken as a measure of the quality of the statistic  $  \theta  ^  \star  $
as an estimator of  $  \theta $
given the loss function  $  w $. 
A partial order relation is thereby introduced on the set of estimators: An estimator  $  T _ {1} $
is preferable to an estimator  $  T _ {2} $
if  $  R _ {w} ( T _ {1} ;  \theta ) \leq  R _ {w} ( T _ {2} ;  \theta ) $. 
In particular, an estimator  $  T $
of the parameter  $  \theta $
is said to be inadmissible (in relation to the loss function  $  w $) 
if an estimator  $  T  ^  \prime  $
exists such that  $  R _ {w} ( T  ^  \prime  ;  \theta ) \leq  R _ {w} ( T;  \theta ) $
for all  $  \theta \in \Theta $, 
and for some  $  \theta $
strict inequality occurs. In this method of comparing the quality of estimators, many estimators prove to be incomparable, and, moreover, the choice of a loss function is to a large extent arbitrary.

It is sometimes possible to find estimators that are optimal within a certain narrower class of estimators. Unbiased estimators form an important class. If the initial experiment is invariant relative to a certain group of transformations, it is natural to restrict to estimators that do not disrupt the symmetry of the problem (see [[Equivariant estimator|Equivariant estimator]]).

Estimators can be compared by their behaviour at  "worst"  points: An estimator  $  T _ {0} $
of  $  \theta $
is called a [[Minimax estimator|minimax estimator]] relative to the loss function  $  w $
if

$$ 
\sup _  \theta   R _ {w} ( T _ {0} ;  \theta )  = \ 
\inf _ { T }  \sup _  \theta   R _ {w} ( T;  \theta ) ,
$$

where the lower bound is taken over all estimators  $  T = T( X) $.

In the Bayesian formulation of the problem (cf. [[Bayesian approach|Bayesian approach]]), the unknown parameter is considered to represent values of the random variable with [[A priori distribution|a priori distribution]]  $  Q $
on  $  \Theta $. 
In this case, the best estimator  $  T _ {0} $
relative to the loss function  $  w $
is defined by the relation

$$ 
r _ {w} ( T _ {0} )  = \ 
{\mathsf E} _ {w} ( T _ {0} ;  \theta )  = \ 
\int\limits _  \Theta  {\mathsf E} _  \theta  w( T _ {0} ;  \theta ) Q( d \theta ) =
$$

$$ 
= \ 
\inf _ { T }  \int\limits _  \Theta  {\mathsf E} _  \theta  w( T;  \theta ) Q( d \theta ) ,
$$

and the lower bound is taken over all estimators  $  T = T( X) $.

There is a distinction between parametric estimation problems, in which  $  \Theta $
is a subset of a finite-dimensional Euclidean space, and non-parametric problems. In parametric problems one usually considers loss functions in the form  $  l( | \theta _ {1} - \theta _ {2} | ) $, 
where  $  l $
is a non-negative, non-decreasing function on  $  \mathbf R  ^ {+} $. 
The most frequently used quadratic loss function  $  | \theta _ {1} - \theta _ {2} |  ^ {2} $
plays an important part.

If  $  T = T( X) $
is a [[Sufficient statistic|sufficient statistic]] for the family  $  \{ { {\mathcal P} _  \theta  } : {\theta \in \Theta } \} $, 
then it is often possible to restrict to estimators  $  \theta  ^  \star  = h( T) $. 
Thus, if  $  \Theta \in \mathbf R  ^ {k} $, 
$  w( \theta _ {1} ;  \theta _ {2} ) = l( | \theta _ {1} - \theta _ {2} | ) $, 
where  $  l $
is a convex function and  $  \theta  ^  \star  $
is any estimator for  $  \theta $, 
an estimator  $  h( T) $
exists that is not worse than  $  \theta  ^  \star  $; 
if  $  \theta  ^  \star  $
is unbiased,  $  h( T) $
can also be chosen unbiased (Blackwell's theorem). If  $  T $
is a complete sufficient statistic for the family  $  \{ {\mathcal P} _  \theta  \} $
and  $  \theta  ^  \star  $
is an unbiased estimator for  $  g( \theta ) $, 
then an unbiased estimator in the form  $  h( T) $
with minimum variance in the class of unbiased estimators exists (the Lehmann–Scheffé theorem).

As a rule, it is assumed that in parametric estimation problems the elements of the family  $  \{ { {\mathcal P} _  \theta  } : {\theta \in \Theta } \} $
are absolutely continuous with respect to a certain  $  \sigma $-
finite measure  $  \mu $
and that the density  $  d {\mathcal P} _  \theta  /d \mu = p( x;  \theta ) $
exists. If  $  p( x;  \theta ) $
is a sufficiently-smooth function of  $  \theta $
and the Fisher information matrix

$$ 
I( \theta )  = \ 
\int\limits _ { \mathfrak X } 
\frac{dp}{d \theta }
 ( x, \theta ) \left ( 
\frac{dp}{d \theta }
 ( x,\ 
\theta ) \right )  ^ {T} 
\frac{\mu ( dx) }{p( x;  \theta ) }

$$

exists, the estimation problem is said to be regular. For regular problems, the accuracy of the estimation is bounded from below by the Cramér–Rao inequality: If  $  \Theta \subset  \mathbf R  ^ {1} $, 
then for any estimator  $  T $,

$$ 
{\mathsf E} _  \theta  | T- \theta |  ^ {2}  \geq  \ 

\frac{( 1+ ( db / {d \theta } ) ( \theta ))  ^ {2} }{I( \theta ) }
 +
b  ^ {2} ( \theta ) ,\ \ 
b( \theta )  =  {\mathsf E} _  \theta  T- \theta .
$$

===Examples of estimation problems 2.===
The most widespread formulation is that in which a sample of size  $  n $
is observed:  $  X _ {1} \dots X _ {n} $
are independent identically-distributed variables taking values in a measurable space  $  ( \mathfrak X , \mathfrak A) $
with common distribution density  $  f( x, \theta ) $
relative to a measure  $  \nu $, 
and  $  \theta \in \Theta $. 
In regular problems, if  $  I( \theta ) $
is the Fisher information on one observation, then the Fisher information of the whole sample  $  I _ {n} ( \theta ) = nI( \theta ) $. 
The Cramér–Rao inequality takes the form

$$ 
{\mathsf E} _  \theta  | T- \theta |  ^ {2}  \geq  \ 

\frac{( 1+ ( db / {d \theta } )( \theta ))  ^ {2} }{nI( \theta ) }
 + b  ^ {2} ( \theta ),\ \ 
$$

$$ 
T  =  T( X _ {1} \dots X _ {n} ).
$$

$  2.1 $. 
Let  $  X _ {j} $
be normal random variables with distribution density

$$ 

\frac{1}{\sqrt {2 \pi } }
  \mathop{\rm exp}
\left \{ - 
\frac{( x- a)  ^ {2} }{2 \sigma  ^ {2} }
 \right \} .
$$

Let the unknown parameter be  $  \theta = ( a, \sigma  ^ {2} ) $; 
$  \overline{X}\; $
and  $  s  ^ {2} $
can serve as estimators for  $  a $
and  $  \sigma  ^ {2} $, 
and  $  ( \overline{X}\; , s  ^ {2} ) $
is then a sufficient statistic. The estimator  $  \overline{X}\; $
is unbiased, while  $  s  ^ {2} $
is biased. If  $  \sigma  ^ {2} $
is known,  $  \overline{X}\; $
is an unbiased estimator of minimal variance, and is a minimax estimator relative to the quadratic loss function.

$  2.2 $. 
Let  $  X _ {j} $
be normal random variables in  $  \mathbf R  ^ {k} $
with density

$$ 

\frac{1}{( 2 \pi )  ^ {k/2} }
  \mathop{\rm exp} \left \{ 
\frac{| x- \theta |  ^ {2} }{2}
 \right \} ,
\  \theta \in \mathbf R  ^ {k} .
$$

The statistic  $  \overline{X}\; $
is an unbiased estimator of  $  \theta $; 
if  $  k \leq  2 $, 
it is admissible relative to the quadratic loss function, if  $  k > 2 $, 
it is inadmissible.

$  2.3 $. 
Let  $  X _ {j} $
be random variables in  $  \mathbf R  ^ {1} $
with unknown distribution density  $  f $
belonging to a given family  $  F $
of densities. For a sufficiently broad class  $  F $, 
this is a non-parametric problem. The problem of estimating  $  f( x _ {0} ) $
at a point  $  x _ {0} $
is a problem of estimating the functional  $  g( f) = f( x _ {0} ) $.

===Example 3.===
The linear regression model. The variables

$$ 
X _ {i}  =  \sum _ {\alpha = 1 } ^ { p }  a _ {\alpha i }  \theta _  \alpha  + \xi _ {i}  $$

are observed; the  $  \xi _ {i} $
are random disturbances,  $  i = 1 \dots n $; 
the matrix  $  \| a _ {\alpha i }  \| $
is known; and the parameter  $  ( \theta _ {1} \dots \theta _ {p} ) $
must be estimated.

===Example 4.===
A segment of a stationary [[Gaussian process|Gaussian process]]  $  x( t) $, 
$  0 \leq  t \leq  T $, 
with rational spectral density  $  | \sum _ {j=} 0  ^ {m} a _ {j} \lambda  ^ {j} |  ^ {2} \cdot  | \sum _ {j=} 0  ^ {n} b _ {j} \lambda  ^ {j} |  ^ {-} 2 $
is observed; the unknown parameters  $  \{ a _ {j} \} $, 
$  \{ b _ {j} \} $
are to be estimated.

==Methods of producing estimators.==
The most widely used [[Maximum-likelihood method|maximum-likelihood method]] recommends that the estimator  $  \widehat \theta   ( X) $
defined as the maximum point of the random function  $  p( X;  \theta ) $
is taken, the so-called maximum-likelihood estimator. If  $  \Theta \subset  \mathbf R  ^ {k} $, 
the maximum-likelihood estimators are to be found among the roots of the likelihood equation

$$ 

\frac{d}{d \theta }
  \mathop{\rm ln}  p( \theta ;  X)  =  0.
$$

In example 3, the method of least squares (cf. [[Least squares, method of|Least squares, method of]]) recommends that the minimum point of the function

$$ 
m( \theta )  =  \sum _ { i= } 1 ^ { n }  \left ( X _ {i} - \sum _  \alpha  a _ {\alpha i } 
\theta _  \alpha  \right )  ^ {2}
$$

be used as the estimator.

Another method is to take a Bayesian estimator  $  T $
relative to a loss function  $  w $
and an a priori distribution  $  Q $, 
although the initial formulation is not Bayesian. For example, if  $  \Theta = \mathbf R  ^ {k} $, 
it is possible to estimate  $  \theta $
by means of

$$ 

\frac{\int\limits _ {- \infty } ^  \infty   \theta p ( X;  \theta )  d \theta }{\int\limits _ {-
\infty } ^  \infty   p( X;  \theta )  d \theta }
 .
$$

This is a Bayesian estimator relative to the quadratic loss function and a uniform a priori distribution.

The method of moments (cf. [[Moments, method of (in probability theory)|Moments, method of (in probability theory)]]) consists of the following. Let  $  \Theta \subset  \mathbf R  ^ {k} $, 
and suppose that there are  $  k $"
good"  estimators  $  a _ {1} ( X) \dots a _ {k} ( X) $
for  $  \alpha _ {1} ( \theta ) \dots \alpha _ {k} ( \theta ) $. 
Estimators by the method of moments are solutions of the system  $  \alpha _ {i} ( \theta ) = a _ {i} $. 
Empirical moments are frequently chosen in the capacity of  $  a _ {i} $(
see example 1).

If the sample  $  X _ {1} \dots X _ {n} $
is observed, then (see example 1) as an estimator for  $  g( {\mathcal P}) $
it is possible to choose  $  g( {\mathcal P} _ {n}  ^  \star  ) $. 
If the function  $  g( {\mathcal P} _ {n}  ^  \star  ) $
is not defined (for example,  $  g( {\mathcal P}) = ( d {\mathcal P} /d \lambda )( x) $, 
where  $  \lambda $
is Lebesgue measure), appropriate modifications  $  g _ {n} ( {\mathcal P} _ {n}  ^  \star  ) $
are chosen. For example, for an estimator of the density a [[Histogram|histogram]] or an estimator of the form

$$ 
\int\limits \phi _ {n} ( x- y)  d {\mathcal P} _ {n}  ^  \star  ( y)
$$

is used.

==Asymptotic behaviour of estimators.==
For the sake of being explicit a problem such as Example 2 is examined, in which  $  \Theta \subset  \mathbf R  ^ {k} $. 
It is to be expected that when  $  n \rightarrow \infty $,  
"good"  estimators will get infinitely close to the characteristic being estimated. A sequence of estimators  $  \theta _ {n}  ^  \star  ( X _ {1} \dots X _ {n} ) $
is called a consistent sequence of estimators of  $  \theta $
if  $  \theta _ {n}  ^  \star  \rightarrow \theta $
in the probability  $  P _  \theta  $
for all  $  \theta $. 
The above methods of producing estimators lead, under broad hypotheses, to consistent estimators (cf. [[Consistent estimator|Consistent estimator]]). The estimators in example 1 are consistent. For regular estimation problems, maximum-likelihood estimators and Bayesian estimators are asymptotically normal with mean  $  \theta $
and correlation matrix  $  ( nI( \theta ))  ^ {-} 1 $. 
Under such conditions, these estimators are asymptotically locally minimax relative to a broad class of loss functions, and they can be considered as being asymptotically optimal (see [[Asymptotically-efficient estimator|Asymptotically-efficient estimator]]).

==Interval estimation.==
A random subset  $  E = E( X) $
of the set  $  \Theta $
is called a confidence region for the estimator  $  \theta $
with confidence coefficient  $  \gamma $
if  $  P _  \theta  \{ E \supset \theta \} = \gamma $(
$  \geq  \gamma $). 
Many confidence regions with a given  $  \gamma $
usually exist, and the problem is to choose the one possessing certain optimal properties (for example, the interval of minimum length, if  $  \Theta \subset  \mathbf R  ^ {1} $). 
Under the conditions of example 2.1, let  $  \sigma = 1 $. 
Then the interval

$$ 
\left [ \overline{X}\; - 
\frac \lambda {\sqrt n }
 , \overline{X}\; + 
\frac \lambda {\sqrt n }

\right ] ,\ \ 
1 - \gamma  =  \sqrt {
\frac{2} \pi 
 } \int\limits _  \lambda  ^  \infty    \mathop{\rm exp} \left \{ - 
\frac{u  ^ {2} }{2}
 \right \}  du ,
$$

is a confidence interval with confidence coefficient  $  \gamma $(
see [[Interval estimator|Interval estimator]]).

====References====
<table><TR><TD valign="top">[1]</TD> <TD valign="top">  R.A. Fisher,   "On the mathematical foundations of theoretical statistics"  ''Phil. Trans. Roy. Soc. London Ser. A'' , '''222'''  (1922)  pp. 309–368</TD></TR><TR><TD valign="top">[2]</TD> <TD valign="top">  A.N. Kolmogorov,   "Sur l'estimation statistique des paramètres de la loi de Gauss"  ''Izv. Akad. Nauk SSSR Ser. Mat.'' , '''6''' :  1  (1942)  pp. 3–32</TD></TR><TR><TD valign="top">[3]</TD> <TD valign="top">  H. Cramér,   "Mathematical methods of statistics" , Princeton Univ. Press  (1946)</TD></TR><TR><TD valign="top">[4]</TD> <TD valign="top">  M.G. Kendall,   A. Stuart,   "The advanced theory of statistics" , '''2. Inference and relationship''' , Griffin  (1979)</TD></TR><TR><TD valign="top">[5]</TD> <TD valign="top">  I.A. Ibragimov,   R.Z. [R.Z. Khas'minskii] Has'minskii,   "Statistical estimation: asymptotic theory" , Springer  (1981)  (Translated from Russian)</TD></TR><TR><TD valign="top">[6]</TD> <TD valign="top">  N.N. Chentsov,   "Statistical decision laws and optimal inference" , Amer. Math. Soc.  (1982)  (Translated from Russian)</TD></TR><TR><TD valign="top">[7]</TD> <TD valign="top">  S. Zacks,   "The theory of statistical inference" , Wiley  (1975)</TD></TR><TR><TD valign="top">[8]</TD> <TD valign="top">  U. Grenander,   "Abstract inference" , Wiley  (1981)</TD></TR></table>

====Comments====

====References====
<table><TR><TD valign="top">[a1]</TD> <TD valign="top">  E.L. Lehmann,   "Theory of point estimation" , Wiley  (1986)</TD></TR></table>