Ronald Fisher
Francis Ysidro Edgeworth
In mathematical statistics, the Fisher information (sometimes simply called information^{[1]}) is a way of measuring the amount of information that an observable random variable X carries about an unknown parameter θ of a distribution that models X. Formally, it is the variance of the score, or the expected value of the observed information. In Bayesian statistics, the asymptotic distribution of the posterior mode depends on the Fisher information and not on the prior (according to the Bernstein–von Mises theorem, which was anticipated by Laplace for exponential families).^{[2]} The role of the Fisher information in the asymptotic theory of maximumlikelihood estimation was emphasized by the statistician Ronald Fisher (following some initial results by Francis Ysidro Edgeworth). The Fisher information is also used in the calculation of the Jeffreys prior, which is used in Bayesian statistics.
The Fisherinformation matrix is used to calculate the covariance matrices associated with maximumlikelihood estimates. It can also be used in the formulation of test statistics, such as the Wald test.
Statistical systems of a scientific nature (physical, biological, etc.) whose likelihood functions obey shift invariance have been shown to obey maximum Fisher information.^{[3]} The level of the maximum depends upon the nature of the system constraints.
Contents

Definition 1

Informal derivation of the Cramér–Rao bound 1.1

Singleparameter Bernoulli experiment 1.2

Matrix form 2

Orthogonal parameters 2.1

Multivariate normal distribution 2.2

Properties 3

Applications 4

Optimal design of experiments 4.1

Jeffreys prior in Bayesian statistics 4.2

Computational neuroscience 4.3

Relation to relative entropy 5

History 6

See also 7

Notes 8

References 9
Definition
The Fisher information is a way of measuring the amount of information that an observable random variable X carries about an unknown parameter θ upon which the probability of X depends. The probability function for X, which is also the likelihood function for θ, is a function f(X; θ); it is the probability mass (or probability density) of the random variable X conditional on the value of θ. The partial derivative with respect to θ of the natural logarithm of the likelihood function is called the score.
Under certain regularity conditions,^{[4]} it can be shown that the first moment of the score (that is, its expected value) is 0:

\operatorname{E} \left[\left. \frac{\partial}{\partial\theta} \log f(X;\theta)\right\theta \right] = \operatorname{E} \left[\left. \frac{\frac{\partial}{\partial\theta} f(X;\theta)}{f(X; \theta)}\right\theta \right] = \int \frac{\frac{\partial}{\partial\theta} f(x;\theta)}{f(x; \theta)} f(x;\theta)\; \mathrm{d}x =

= \int \frac{\partial}{\partial\theta} f(x;\theta)\; \mathrm{d}x = \frac{\partial}{\partial\theta} \int f(x; \theta)\; \mathrm{d}x = \frac{\partial}{\partial\theta} \; 1 = 0.
The second moment is called the Fisher information:

\mathcal{I}(\theta)=\operatorname{E} \left[\left. \left(\frac{\partial}{\partial\theta} \log f(X;\theta)\right)^2\right\theta \right] = \int \left(\frac{\partial}{\partial\theta} \log f(x;\theta)\right)^2 f(x; \theta)\; \mathrm{d}x\,,
where, for any given value of θ, the expression E[...θ] denotes the conditional expectation over values for X with respect to the probability function f(x; θ) given θ. Note that 0 \leq \mathcal{I}(\theta) < \infty. A random variable carrying high Fisher information implies that the absolute value of the score is often high. The Fisher information is not a function of a particular observation, as the random variable X has been averaged out.
Since the expectation of the score is zero, the Fisher information is also the variance of the score.
If log f(x; θ) is twice differentiable with respect to θ, and under certain regularity conditions, then the Fisher information may also be written as^{[5]}

\mathcal{I}(\theta) =  \operatorname{E} \left[\left. \frac{\partial^2}{\partial\theta^2} \log f(X;\theta)\right\theta \right]\,,
since

\frac{\partial^2}{\partial\theta^2} \log f(X;\theta) = \frac{\frac{\partial^2}{\partial\theta^2} f(X;\theta)}{f(X; \theta)} \;\; \left( \frac{\frac{\partial}{\partial\theta} f(X;\theta)}{f(X; \theta)} \right)^2 = \frac{\frac{\partial^2}{\partial\theta^2} f(X;\theta)}{f(X; \theta)} \;\; \left( \frac{\partial}{\partial\theta} \log f(X;\theta)\right)^2
and

\operatorname{E} \left[\left. \frac{\frac{\partial^2}{\partial\theta^2} f(X;\theta)}{f(X; \theta)}\right\theta \right] = \cdots = \frac{\partial^2}{\partial\theta^2} \int f(x; \theta)\; \mathrm{d}x = \frac{\partial^2}{\partial\theta^2} \; 1 = 0.
Thus, the Fisher information is the negative of the expectation of the second derivative with respect to θ of the natural logarithm of f. Information may be seen to be a measure of the "curvature" of the support curve near the maximum likelihood estimate of θ. A "blunt" support curve (one with a shallow maximum) would have a low negative expected second derivative, and thus low information; while a sharp one would have a high negative expected second derivative and thus high information.
Information is additive, in that the information yielded by two independent experiments is the sum of the information from each experiment separately:

\mathcal{I}_{X,Y}(\theta) = \mathcal{I}_X(\theta) + \mathcal{I}_Y(\theta).
This result follows from the elementary fact that if random variables are independent, the variance of their sum is the sum of their variances. In particular, the information in a random sample of size n is n times that in a sample of size 1, when observations are independent and identically distributed.
The information provided by a sufficient statistic is the same as that of the sample X. This may be seen by using Neyman's factorization criterion for a sufficient statistic. If T(X) is sufficient for θ, then

f(X;\theta) = g(T(X), \theta) h(X) \!
for some functions g and h. See sufficient statistic for a more detailed explanation. The equality of information then follows from the following fact:

\frac{\partial}{\partial\theta} \log \left[f(X ;\theta)\right] = \frac{\partial}{\partial\theta} \log \left[g(T(X);\theta)\right]
which follows from the definition of Fisher information, and the independence of h(X) from θ. More generally, if T = t(X) is a statistic, then

\mathcal{I}_T(\theta) \leq \mathcal{I}_X(\theta)
with equality if and only if T is a sufficient statistic.
Informal derivation of the Cramér–Rao bound
The Cramér–Rao bound states that the inverse of the Fisher information is a lower bound on the variance of any unbiased estimator of θ. H.L. Van Trees (1968) and B. Roy Frieden (2004) provide the following method of deriving the Cramér–Rao bound, a result which describes use of the Fisher information, informally:
Consider an unbiased estimator \hat\theta(X). Mathematically, we write

\operatorname{E}\left[ \left. \hat\theta(X)  \theta \right \theta \right] = \int \left[ \hat\theta(x)  \theta \right] \cdot f(x ;\theta) \, \mathrm{d}x = 0.
The likelihood function f(X; θ) describes the probability that we observe a given sample x given a known value of θ. If f is sharply peaked with respect to changes in θ, it is easy to intuit the "correct" value of θ given the data, and hence the data contains a lot of information about the parameter. If the likelihood f is flat and spreadout, then it would take many, many samples of X to estimate the actual "true" value of θ. Therefore, we would intuit that the data contain much less information about the parameter.
Now, we use the product rule to differentiate the unbiasedness condition above to get

\frac{\partial}{\partial\theta} \int \left[ \hat\theta(x)  \theta \right] \cdot f(x ;\theta) \, \mathrm{d}x = \int \left(\hat\theta(x)\theta\right) \frac{\partial f}{\partial\theta} \, \mathrm{d}x  \int f \, \mathrm{d}x = 0.
We now make use of two facts. The first is that the likelihood f is just the probability of the data given the parameter. Since it is a probability, it must be normalized, implying that

\int f \, \mathrm{d}x = 1.
Second, we know from basic calculus that

\frac{\partial f}{\partial\theta} = f \, \frac{\partial \log f}{\partial\theta}.
Using these two facts in the above let us write

\int \left(\hat\theta\theta\right) f \, \frac{\partial \log f}{\partial\theta} \, \mathrm{d}x = 1.
Factoring the integrand gives

\int \left(\left(\hat\theta\theta\right) \sqrt{f} \right) \left( \sqrt{f} \, \frac{\partial \log f}{\partial\theta} \right) \, \mathrm{d}x = 1.
If we square the equation, the Cauchy–Schwarz inequality lets us write

\left[ \int \left(\hat\theta  \theta\right)^2 f \, \mathrm{d}x \right] \cdot \left[ \int \left( \frac{\partial \log f}{\partial\theta} \right)^2 f \, \mathrm{d}x \right] \geq 1.
The rightmost factor is defined to be the Fisher Information

\mathcal{I}\left(\theta\right) = \int \left( \frac{\partial \log f}{\partial\theta} \right)^2 f \, \mathrm{d}x.
The leftmost factor is the expected meansquared error of the estimator θ^{^}, since

\operatorname{E}\left[ \left. \left( \hat\theta\left(X\right)  \theta \right)^2 \right \theta \right] = \int \left(\hat\theta  \theta\right)^2 f \, \mathrm{d}x.
Notice that the inequality tells us that, fundamentally,

\operatorname{Var}\left(\hat\theta\right) \, \geq \, \frac{1}{\mathcal{I}\left(\theta\right)}.
In other words, the precision to which we can estimate θ is fundamentally limited by the Fisher Information of the likelihood function.
Singleparameter Bernoulli experiment
A Bernoulli trial is a random variable with two possible outcomes, "success" and "failure", with success having a probability of θ. The outcome can be thought of as determined by a coin toss, with the probability of heads being θ and the probability of tails being 1 − θ.
The Fisher information contained in n independent Bernoulli trials may be calculated as follows. In the following, A represents the number of successes, B the number of failures, and n = A + B is the total number of trials.
\begin{align} \mathcal{I}(\theta) & = \operatorname{E} \left[ \left. \frac{\partial^2}{\partial\theta^2} \log(f(A;\theta)) \right \theta \right] \qquad (1) \\ & = \operatorname{E} \left[ \left. \frac{\partial^2}{\partial\theta^2} \log \left( \theta^A(1\theta)^B\frac{(A+B)!}{A!B!} \right) \right \theta \right] \qquad (2) \\ & = \operatorname{E} \left[ \left. \frac{\partial^2}{\partial\theta^2} \left( A \log (\theta) + B \log(1\theta) \right) \right \theta \right] \qquad (3) \\ & = \operatorname{E} \left[ \left. \frac{\partial}{\partial\theta} \left( \frac{A}{\theta}  \frac{B}{1\theta} \right) \right \theta \right] \qquad (4) \\ & = +\operatorname{E} \left[ \left. \frac{A}{\theta^2} + \frac{B}{(1\theta)^2} \right \theta \right] \qquad (5) \\ & = \frac{n\theta}{\theta^2} + \frac{n(1\theta)}{(1\theta)^2} \qquad (6) \\ & \text{since the expected value of }A\text{ given }\theta\text{ is }n\theta,\text{ etc.} \\ & = \frac{n}{\theta(1\theta)} \qquad (7) \end{align}
(1) defines Fisher information. (2) invokes the fact that the information in a sufficient statistic is the same as that of the sample itself. (3) expands the natural logarithm term and drops a constant. (4) and (5) differentiate with respect to θ. (6) replaces A and B with their expectations. (7) is algebra.
The end result, namely,

\mathcal{I}(\theta) = \frac{n}{\theta(1\theta)},
is the reciprocal of the variance of the mean number of successes in n Bernoulli trials, as expected (see last sentence of the preceding section).
Matrix form
When there are N parameters, so that θ is a N × 1 vector \theta = \begin{bmatrix} \theta_{1}, \theta_{2}, \dots , \theta_{N} \end{bmatrix}^{\mathrm T}, then the Fisher information takes the form of an N × N matrix, the Fisher Information Matrix (FIM), with typical element

{\left(\mathcal{I} \left(\theta \right) \right)}_{i, j} = \operatorname{E} \left[\left. \left(\frac{\partial}{\partial\theta_i} \log f(X;\theta)\right) \left(\frac{\partial}{\partial\theta_j} \log f(X;\theta)\right) \right\theta\right].
The FIM is a N × N positive semidefinite symmetric matrix, defining a Riemannian metric on the Ndimensional parameter space, thus connecting Fisher information to differential geometry. In that context, this metric is known as the Fisher information metric, and the topic is called information geometry.
Under certain regularity conditions, the Fisher Information Matrix may also be written as

{\left(\mathcal{I} \left(\theta \right) \right)}_{i, j} =  \operatorname{E} \left[\left. \frac{\partial^2}{\partial\theta_i \, \partial\theta_j} \log f(X;\theta) \right\theta\right]\,.
The metric is interesting in several ways; it can be derived as the Hessian of the relative entropy; it can be understood as a metric induced from the Euclidean metric, after appropriate change of variable; in its complexvalued form, it is the Fubini–Study metric.
Orthogonal parameters
We say that two parameters θ_{i} and θ_{j} are orthogonal if the element of the ith row and jth column of the Fisher information matrix is zero. Orthogonal parameters are easy to deal with in the sense that their maximum likelihood estimates are independent and can be calculated separately. When dealing with research problems, it is very common for the researcher to invest some time searching for an orthogonal parametrization of the densities involved in the problem.
Multivariate normal distribution
The FIM for a Nvariate multivariate normal distribution has a special form. Let \mu(\theta) = \begin{bmatrix} \mu_{1}(\theta), \mu_{2}(\theta), \dots , \mu_{N}(\theta) \end{bmatrix}^\mathrm{T}, and let Σ(θ) be the covariance matrix. Then the typical element \mathcal{I}_{m,n}, 0 ≤ m, n < K, of the FIM for X ∼ N(μ(θ), Σ(θ)) is:

\mathcal{I}_{m,n} = \frac{\partial \mu^\mathrm{T}}{\partial \theta_m} \Sigma^{1} \frac{\partial \mu}{\partial \theta_n} + \frac{1}{2} \operatorname{tr} \left( \Sigma^{1} \frac{\partial \Sigma}{\partial \theta_m} \Sigma^{1} \frac{\partial \Sigma}{\partial \theta_n} \right),
where (..)^\mathrm{T} denotes the transpose of a vector, tr(..) denotes the trace of a square matrix, and:

\frac{\partial \mu}{\partial \theta_m} = \begin{bmatrix} \frac{\partial \mu_1}{\partial \theta_m} & \frac{\partial \mu_2}{\partial \theta_m} & \cdots & \frac{\partial \mu_N}{\partial \theta_m} \end{bmatrix}^\mathrm{T};

\frac{\partial \Sigma}{\partial \theta_m} = \begin{bmatrix} \frac{\partial \Sigma_{1,1}}{\partial \theta_m} & \frac{\partial \Sigma_{1,2}}{\partial \theta_m} & \cdots & \frac{\partial \Sigma_{1,N}}{\partial \theta_m} \\ \\ \frac{\partial \Sigma_{2,1}}{\partial \theta_m} & \frac{\partial \Sigma_{2,2}}{\partial \theta_m} & \cdots & \frac{\partial \Sigma_{2,N}}{\partial \theta_m} \\ \\ \vdots & \vdots & \ddots & \vdots \\ \\ \frac{\partial \Sigma_{N,1}}{\partial \theta_m} & \frac{\partial \Sigma_{N,2}}{\partial \theta_m} & \cdots & \frac{\partial \Sigma_{N,N}}{\partial \theta_m} \end{bmatrix}.
Note that a special, but very common, case is the one where Σ(θ) = Σ, a constant. Then

\mathcal{I}_{m,n} = \frac{\partial \mu^\mathrm{T}}{\partial \theta_m} \Sigma^{1} \frac{\partial \mu}{\partial \theta_n}.\
In this case the Fisher information matrix may be identified with the coefficient matrix of the normal equations of least squares estimation theory.
Another special case is that the mean and covariance depends on two different vector parameters, say, β and θ. This is especially popular in the analysis of spatial data, which uses a linear model with correlated residuals. We have
\mathcal{I}\left( \beta ,\theta \right)=\text{diag}\left( \mathcal{I}\left( \beta \right),\mathcal{I}\left( \theta \right) \right)
where
\mathcal{I}=\frac{\partial }}{\partial }\frac{\partial \mu }{\partial },
\mathcal{I}=\frac{1}{2}\operatorname{tr}\left( \frac{\partial \Sigma }{\partial }\frac{\partial \Sigma }{\partial } \right)
The proof of this special case is given in literature.^{[6]} Using the same technique in this paper, it's not difficult to prove the original result.
Properties
Reparametrization
The Fisher information depends on the parametrization of the problem. If θ and η are two scalar parametrizations of an estimation problem, and θ is a continuously differentiable function of η, then

{\mathcal I}_\eta(\eta) = {\mathcal I}_\theta(\theta(\eta)) \left( \frac_{\text{Fisher info.}}(\theta'\theta)+\cdots
Thus the Fisher information represents the curvature of the relative entropy.
Schervish (1995: §2.3) says the following.
One advantage KullbackLeibler information has over Fisher information is that it is not affected by changes in parameterization. Another advantage is that KullbackLeibler information can be used even if the distributions under consideration are not all members of a parametric family. ...
Another advantage to KullbackLeibler information is that no smoothness conditions on the densities … are needed.
History
The Fisher information was discussed by several early statisticians, notably F. Y. Edgeworth.^{[14]} For example, Savage^{[15]} says: "In it [Fisher information], he [Fisher] was to some extent anticipated (Edgeworth 1908–9 esp. 502, 507–8, 662, 677–8, 82–5 and references he [Edgeworth] cites including Pearson and Filon 1898 [. . .])." There are a number of early historical sources^{[16]} and a number of reviews of this early work.^{[17]}^{[18]}^{[19]}
See also
Other measures employed in information theory:
Notes

^ Lehmann & Casella, p. 115

^ Lucien Le Cam (1986) Asymptotic Methods in Statistical Decision Theory: Pages 336 and 618–621 (von Mises and Bernstein).

^ Frieden & Gatenby (2013)

^ Suba Rao. "Lectures on statistical inference" (PDF).

^ Lehmann & Casella, eq. (2.5.16).

^ Mardia, K. V.; Marshall, R. J. (1984). "Maximum likelihood estimation of models for residual covariance in spatial regression".

^ Lehmann & Casella, eq. (2.5.11).

^ Lehmann & Casella, eq. (2.6.16)

^ Janke, W.; Johnston, D. A.; Kenna, R. (2004). "Information Geometry and Phase Transitions". Physica A 336 (1–2): 181.

^ Prokopenko, M.; Lizier, Joseph T.; Lizier, J. T.; Obst, O.; Wang, X. R. (2011). "Relating Fisher information to order parameters". Physical Review E 84 (4): 041116.

^ Pukelsheim, Friedrick (1993). Optimal Design of Experiments. New York: Wiley.

^ Bernardo, Jose M.; Smith, Adrian F. M. (1994). Bayesian Theory. New York: John Wiley & Sons.

^ Gourieroux & Montfort (1995), page 87

^ Savage (1976)

^ Savage(1976), page 156

^ Edgeworth (September 1908, December 1908)

^ Pratt (1976)

^ Stigler (1978, 1986, 1999)

^ Hald (1998, 1999)
References




Frieden, B. R. (2004) Science from Fisher Information: A Unification. Cambridge Univ. Press. ISBN 0521009111.

Frieden, B. Roy; Gatenby, Robert A. (2013). "Principle of maximum Fisher information from Hardy's axioms applied to statistical systems". Physical Review E 88 (4).

Hald, A. (May 1999). "On the History of Maximum Likelihood in Relation to Inverse Probability and Least Squares".

Hald, A. (1998). A History of Mathematical Statistics from 1750 to 1930. New York: Wiley.



Pratt, John W. (May 1976). "F. Y. Edgeworth and R. A. Fisher on the Efficiency of Maximum Likelihood Estimation".


Schervish, Mark J. (1995). Theory of Statistics. New York: Springer.




Van Trees, H. L. (1968). Detection, Estimation, and Modulation Theory, Part I. New York: Wiley.
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