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Probability density function
Gamma distribution pdf
Cumulative distribution function
Gamma distribution cdf
Parameters k > 0\, shape (real)
\theta > 0\, scale (real)
Support x \in [0; \infty)\!
pdf x^{k-1} \frac{\exp\left(-x/\theta\right)}{\Gamma(k)\,\theta^k}
cdf \frac{\gamma(k, x/\theta)}{\Gamma(k)}
Mean k \theta\,
Mode (k-1) \theta\, for k \geq 1\,
Variance k \theta^2\,
Skewness \frac{2}{\sqrt{k}}
Kurtosis \frac{6}{k}
Entropy k\theta+(1-k)\ln(\theta)+\ln(\Gamma(k))\,
mgf (1 - \theta\,t)^{-k} for t < 1/\theta
Char. func. (1 - \theta\,i\,t)^{-k}

In probability theory and statistics, the gamma distribution is a continuous probability distribution. For integer values of the parameter k it is also known as the Erlang distribution.

Probability density functionEdit

The probability density function of the gamma distribution can be expressed in terms of the gamma function:

 f(x;k,\theta) = x^{k-1} \frac{e^{-x/\theta}}{\theta^k \, \Gamma(k)} 
 \ \mathrm{for}\ x > 0 \,\!

where k > 0 is the shape parameter and \theta > 0 is the scale parameter of the gamma distribution. (NOTE: this parameterization is what is used in the infobox and the plots.)

Alternatively, the gamma distribution can be parameterized in terms of a shape parameter \alpha = k and an inverse scale parameter \beta = 1/\theta, called a rate parameter:

 g(x;\alpha,\beta) = x^{\alpha-1}  \frac{\beta^{\alpha} \, e^{-\beta\,x} }{\Gamma(\alpha)}  \ \mathrm{for}\ x > 0 \,\!

Both parameterizations are common because they are convenient to use in certain situations and fields.


The cumulative distribution function can be expressed in terms of the incomplete gamma function,

 F(x;k,\theta) = \int_0^x f(u;k,\theta)\,du  
  = \frac{\gamma(k, x/\theta)}{\Gamma(k)} \,\!

The information entropy is given by:


where \psi(k) is the polygamma function.

If X_i \sim \mathrm{Gamma}(\alpha_i, \beta) for i=1, 2, \cdots, N and \bar{\alpha} = \sum_{k=1}^N \alpha_i then

\sum_{i=1}^N X_i
\mathrm{Gamma} \left( \bar{\alpha}, \beta \right)

provided all X_i are independent. The gamma distribution exhibits infinite divisibility.

If X \sim \operatorname{Gamma}(k, \theta), then \frac X \theta \sim \operatorname{Gamma}(k, 1). Or, more generally, for any t > 0 it holds that tX \sim \operatorname{Gamma} (k, t \theta). That is the meaning of θ (or β) being the scale parameter.

Parameter estimation Edit

The likelihood function is

L=\prod_{i=1}^N f(x_i;k,\theta)

from which we calculate the log-likelihood function

\ell=(k-1)\sum_{i=1}^N\ln(x_i)-\sum x_i/\theta-Nk\ln(\theta)-N\ln\Gamma(k)

Finding the maximum with respect to \theta by taking the derivative and setting it equal to zero yields the maximum likelihood estimate of the \theta parameter:

\theta=\frac{1}{kN}\sum_{i=1}^N x_i

Substituting this into the log-likelihood function gives:

\ell=(k-1)\sum_{i=1}^N\ln(x_i)-Nk-Nk\ln\left(\frac{\sum x_i}{kN}\right)-N\ln\Gamma(k)

Finding the maximum with respect to k by taking the derivative and setting it equal to zero yields:

\ln(k)-\psi(k)=\ln\left(\frac{1}{N}\sum_{i=1}^N x_i\right)-\frac{1}{N}\sum_{i=1}^N\ln(x_i)

where \psi(k)=\frac{\Gamma'(k)}{\Gamma(k)} is the digamma function.

There is no closed-form solution for k. The function is numerically very well behaved, so if a numerical solution is desired, it can be found using Newton's method. An initial value of k can be found either using the method of moments, or using the approximation:

\ln(k)-\psi(k) \approx \frac{1}{k}\left(\frac{1}{2} + \frac{1}{12k+2}\right)

If we let s = \ln\left(\frac{1}{N}\sum_{i=1}^N x_i\right)-\frac{1}{N}\sum_{i=1}^N\ln(x_i), then k is approximately

k \approx \frac{3-s+\sqrt{(s-3)^2 + 24s}}{12s}

which is within 1.5% of the correct value.

Generating Gamma random variables Edit

Given the scaling property above, it is enough to generate Gamma variables with \beta = 1 as we can later convert to any value of β with simple division.

Using the fact that if X \sim \operatorname{Gamma}(1, 1), then also X \sim \operatorname {Exp} (1), and the method of generating exponential variables, we conclude that if U is uniformly distributed on (0, 1], then -\ln U \sim \operatorname{Gamma} (1, 1). Now, using the "α-addition" property of Gamma distribution, we expand this result:

\sum _{k=1} ^n {-\ln U_k} \sim \operatorname{Gamma} (n, 1),

where U_k are all uniformly distributed on (0, 1 ] and independent.

All that is left now is to generate a variable distributed as \operatorname{Gamma} (\delta, 1) for 0 < \delta < 1 and apply the "α-addition" property once more. This is the most difficult part, however.

We provide an algorithm without proof. It is an instance of the acceptance-rejection method:

  1. Let m be 1.
  2. Generate V_{2m - 1} and V_{2m} — independent uniformly distributed on (0, 1] variables.
  3. If V_{2m - 1} \le v_0, where v_0 = \frac e {e + \delta}, then go to step 4, else go to step 5.
  4. Let \xi_m = \left( \frac {V_{2m - 1}} {v_0} \right) ^{\frac 1 \delta}, \ \eta_m = V_{2m} \xi _m^ {\delta - 1}. Go to step 6.
  5. Let \xi_m = 1 - \ln {\frac {V_{2m - 1} - v_0} {1 - v_0}}, \ \eta_m = V_{2m} e^{-\xi_m}.
  6. If \eta_m > \xi_m^{\delta - 1} e^{-\xi_m}, then increment m and go to step 2.
  7. Assume \xi = \xi_m to be the realization of \operatorname {Gamma} (\delta, 1).

Now, to summarize,

\frac 1 \beta \left( \xi - \sum _{k=1} ^{[\alpha]} {\ln U_k} \right) \sim \operatorname{Gamma}(\alpha, \beta) ,

where [\alpha] is the integral part of α, ξ has been generating using the algorithm above with \delta = \{\alpha\} (the fractional part of α), U_k and V_l are distributed as explained above and are all independent.

Related distributionsEdit

References Edit

  • R. V. Hogg and A. T. Craig. Introduction to Mathematical Statistics, 4th edition. New York: Macmillan, 1978. (See Section 3.3.)

See also Edit

es:Distribución gammafi:Gamma-jakauma sv:Gammafördelning

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