Moment generating function

Let X be a random variable with CDF F_X. The moment generating function (mgf) of X (or F_X), denoted by M_X(t), is M_X(t) = \operatorname E \left[e^{tX}\right] provided this expectation exists for t in some open neighborhood of 0. That is, there is an h > 0 such that for all t satisfying -h , \operatorname E \left[e^{tX}\right] exists. If the expectation does not exist in an open neighborhood of 0, we say that the moment generating function does not exist. In other words, the moment generating function of is the expectation of the random variable e^{tX}. More generally, when \mathbf X = ( X_1, \ldots, X_n)^{\mathrm{T}}, an n-dimensional random vector, and \mathbf t is a fixed vector, one uses \mathbf t \cdot \mathbf X = \mathbf t^\mathrm T\mathbf X instead of  M_{\mathbf X}(\mathbf t) := \operatorname E \left[e^{\mathbf t^\mathrm T\mathbf X}\right]. M_X(0) always exists and is equal to 1. However, a key problem with moment generating functions is that moments and the moment generating function may not exist, as the integrals need not converge absolutely. By contrast, the characteristic function or Fourier transform always exists (because it is the integral of a bounded function on a space of finite measure), and for some purposes may be used instead. The moment generating function is so named because it can be used to find the moments of the distribution. The series expansion of e^{tX} is e^{t X} = 1 + t X + \frac{t^2 X^2}{2!} + \frac{t^3 X^3}{3!} + \cdots + \frac{t^n X^n}{n!} + \cdots. Hence, \begin{align} M_X(t) &= \operatorname E [e^{t X}] \\[1ex] &= 1 + t \operatorname E[X] + \frac{t^2 \operatorname E[X^2]}{2!} + \frac{t^3 \operatorname E[X^3]}{3!} + \cdots + \frac{t^n\operatorname E [X^n]}{n!}+\cdots \\[1ex] & = 1 + t m_1 + \frac{t^2 m_2}{2!} + \frac{t^3 m_3}{3!} + \cdots + \frac{t^n m_n}{n!} + \cdots, \end{align} where m_n is the moment. Differentiating M_X(t) i times with respect to t and setting t = 0, we obtain the i-th moment about the origin, m_i. If X is a continuous random variable, the following relation between its moment generating function M_X(t) and the two-sided Laplace transform of its probability density function f_X(x) holds: M_X(t) = \mathcal{L}\{f_X\}(-t), since the PDF's two-sided Laplace transform is given as \mathcal{L}\{f_X\}(s) = \int_{-\infty}^\infty e^{-sx} f_X(x)\, dx, and the moment generating function's definition expands (by the law of the unconscious statistician) to M_X(t) = \operatorname E \left[e^{tX}\right] = \int_{-\infty}^\infty e^{tx} f_X(x)\, dx. This is consistent with the characteristic function of X being a Wick rotation of M_X(t) when the moment generating function exists, as the characteristic function of a continuous random variable X is the Fourier transform of its probability density function f_X(x), and in general when a function f(x) is of exponential order, the Fourier transform of f is a Wick rotation of its two-sided Laplace transform in the region of convergence. See the relation of the Fourier and Laplace transforms for further information. ==Examples==

Here are some examples of the moment generating function and the characteristic function for comparison. It can be seen that the characteristic function is a Wick rotation of the moment generating function M_X(t) when the latter exists. \operatorname{MultiCauchy}(\mu, \Sigma) ==Calculation==

The moment generating function is the expectation of a function of the random variable, it can be written as: • For a discrete probability mass function, M_X(t)=\sum_{i=0}^\infty e^{tx_i}\, p_i • For a continuous probability density function, M_X(t) = \int_{-\infty}^\infty e^{tx} f(x)\,dx • In the general case: M_X(t) = \int_{-\infty}^\infty e^{tx}\,dF(x), using the Riemann–Stieltjes integral, and where F is the cumulative distribution function. This is simply the Laplace-Stieltjes transform of F, but with the sign of the argument reversed. Note that for the case where X has a continuous probability density function f(x), M_X(-t) is the two-sided Laplace transform of f(x). \begin{align} M_X(t) & = \int_{-\infty}^\infty e^{tx} f(x)\,dx \\[1ex] & = \int_{-\infty}^\infty \left( 1+ tx + \frac{t^2 x^2}{2!} + \cdots + \frac{t^n x^n}{n!} + \cdots\right) f(x)\,dx \\[1ex] & = 1 + tm_1 + \frac{t^2 m_2}{2!} + \cdots + \frac{t^n m_n}{n!} +\cdots, \end{align} where m_n is the nth moment. Linear transformations of random variables If random variable X has moment generating function M_X(t), then \alpha X + \beta has moment generating function M_{\alpha X + \beta}(t) = e^{\beta t}M_X(\alpha t) M_{\alpha X + \beta}(t) = \operatorname{E}\left[e^{(\alpha X + \beta) t}\right] = e^{\beta t} \operatorname{E}\left[e^{\alpha Xt}\right] = e^{\beta t} M_X(\alpha t) Linear combination of independent random variables If S_n = \sum_{i=1}^n a_i X_i, where the are independent random variables and the are constants, then the probability density function for is the convolution of the probability density functions of each of the , and the moment generating function for is given by M_{S_n}(t) = M_{X_1}(a_1t) M_{X_2}(a_2t) \cdots M_{X_n}(a_nt) \, . X, then G_{S_N}(z) = G_N(G_X(z)). --------> Vector-valued random variables For vector-valued random variables \mathbf X with real components, the moment generating function is given by M_X(\mathbf t) = \operatorname{E}\left[e^{\langle \mathbf t, \mathbf X \rangle}\right] where \mathbf t is a vector and \langle \cdot, \cdot \rangle is the dot product. ==Important properties==

Moment generating functions are positive and log-convex, with M(0) = 1. An important property of the moment generating function is that it uniquely determines the distribution. In other words, if X and Y are two random variables and for all values of , M_X(t) = M_Y(t), then F_X(x) = F_Y(x) for all values of (or equivalently and have the same distribution). This statement is not equivalent to the statement "if two distributions have the same moments, then they are identical at all points." This is because in some cases, the moments exist and yet the moment generating function does not, because the limit \lim_{n \to \infty} \sum_{i=0}^n \frac{t^i m_i}{i!} may not exist. The log-normal distribution is an example of when this occurs: its moments are \operatorname{E}[X^n] = e^{n\mu+n^2\sigma^2/2} and are all finite but its moment generating function \operatorname E \left[e^{tX}\right] is not defined for any positive as the integral diverges and so not in a neighbourhood of 0; there are other distributions with the same moments. Calculations of moments The moment generating function is so called because if it exists on an open interval around , then it is the exponential generating function of the moments of the probability distribution: m_n = \operatorname{E}\left[ X^n \right] = M_X^{(n)}(0) = \left. \frac{d^n M_X}{dt^n}\right|_{t=0}. That is, with being a nonnegative integer, the -th moment about 0 is the -th derivative of the moment generating function, evaluated at . ==Other properties==

Jensen's inequality provides a simple lower bound on the moment generating function: M_X(t) \geq e^{\mu t}, where \mu is the mean of . The moment generating function can be used in conjunction with Markov's inequality to bound the upper tail of a real random variable . This statement is also called the Chernoff bound. Since x \mapsto e^{xt} is monotonically increasing for t>0, we have \Pr(X \ge a) = \Pr(e^{tX} \ge e^{ta}) \le e^{-at} \operatorname{E}\left[e^{tX}\right] = e^{-at}M_X(t) for any t>0 and any , provided M_X(t) exists. For example, when is a standard normal distribution and a > 0, we can choose t=a and recall that M_X(t)=e^{t^2/2}. This gives \Pr(X\ge a)\le e^{-a^2/2}, which is within a factor of of the exact value. Various lemmas, such as Hoeffding's lemma or Bennett's inequality provide bounds on the moment generating function in the case of a zero-mean, bounded random variable. When X is non-negative, the moment generating function gives a simple, useful bound on the moments: \operatorname{E}[X^m] \le \left(\frac{m}{te}\right)^m M_X(t), For any X,m\ge 0 and t>0. This follows from the inequality 1+x\le e^x into which we can substitute x'=tx/m-1 implies tx/m\le e^{tx/m-1} for any Now, if t > 0 and x,m\ge 0, this can be rearranged to x^m \le (m/(te))^m e^{tx}. Taking the expectation on both sides gives the bound on \operatorname{E}[X^m] in terms of \operatorname{E}[e^{tX}]. As an example, consider X\sim\text{Chi-Squared} with k degrees of freedom. Then from the examples M_X(t) = (1-2t)^{-k/2}. Picking t=m/(2m+k) and substituting into the bound: \operatorname{E}[X^m] \le {\left(1 + 2m/k\right)}^{k/2} e^{-m} {\left(k + 2m\right)}^m. We know that in this case the correct bound is \operatorname{E}[X^m]\le 2^m \Gamma(m+k/2)/\Gamma(k/2). To compare the bounds, we can consider the asymptotics for large k. Here the moment generating function bound is k^m(1+m^2/k + O(1/k^2)), where the real bound is k^m(1+(m^2-m)/k + O(1/k^2)). The moment generating function bound is thus very strong in this case. ==Relation to other functions==

Related to the moment generating function are a number of other transforms that are common in probability theory: ;Characteristic function: The characteristic function \varphi_X(t) is related to the moment generating function via \varphi_X(t) = M_{iX}(t) = M_X(it): the characteristic function is the moment generating function of iX or the moment generating function of X evaluated on the imaginary axis. This function can also be viewed as the Fourier transform of the probability density function, which can therefore be deduced from it by inverse Fourier transform. ;Cumulant-generating function: The cumulant-generating function is defined as the logarithm of the moment generating function; some instead define the cumulant-generating function as the logarithm of the characteristic function, while others call this latter the second cumulant-generating function. ;Probability-generating function: The probability-generating function is defined as G(z) = \operatorname{E}\left[z^X\right]. This immediately implies that G(e^t) = \operatorname{E}\left[e^{tX}\right] = M_X(t). ==See also==