# Exponential distribution

Exponential distribution
parameters: Probability density function Cumulative distribution function λ > 0 rate, or inverse scale x ∈ [0, ∞) λ e−λx 1 − e−λx λ−1 λ−1 ln 2 0 λ−2 2 6 1 − ln(λ) $\left(1 - \frac{t}{\lambda}\right)^{-1}\,$ $\left(1 - \frac{it}{\lambda}\right)^{-1}\,$

In probability theory and statistics, the exponential distribution (a.k.a. negative exponential distribution) is a family of continuous probability distributions. It describes the time between events in a Poisson process, i.e. a process in which events occur continuously and independently at a constant average rate.

Note that the exponential distribution is not the same as the class of exponential families of distributions, which is a large class of probability distributions that includes the exponential distribution as one of its members, but also includes the normal distribution, binomial distribution, gamma distribution, Poisson, and many others.

## Characterization

### Probability density function

The probability density function (pdf) of an exponential distribution is

$f(x;\lambda) = \begin{cases} \lambda e^{-\lambda x}, & x \ge 0, \\ 0, & x < 0. \end{cases}$

Alternatively, this can be defined using the Heaviside step function, H(x).

$f(x;\lambda) = \mathrm \lambda e^{-\lambda x} H(x) \!$

Here λ > 0 is the parameter of the distribution, often called the rate parameter. The distribution is supported on the interval [0, ∞). If a random variable X has this distribution, we write X ~ Exp(λ).

The exponential distribution exhibits infinite divisibility.

### Cumulative distribution function

The cumulative distribution function is given by

$F(x;\lambda) = \begin{cases} 1-e^{-\lambda x}, & x \ge 0, \\ 0, & x < 0. \end{cases}$

Alternatively, this can be defined using the Heaviside step functionH(x).

$F(x;\lambda) = \mathrm (1-e^{-\lambda x}) H(x) \!$

### Alternative parameterization

A commonly used alternative parameterization is to define the probability density function (pdf) of an exponential distribution as

$f(x;\beta) = \begin{cases} \frac{1}{\beta} e^{-x/\beta}, & x \ge 0, \\ 0, & x < 0. \end{cases}$

where β > 0 is a scale parameter of the distribution and is the reciprocal of the rate parameter, λ, defined above. In this specification, β is a survival parameter in the sense that if a random variable X is the duration of time that a given biological or mechanical system manages to survive and X ~ Exponential(β) then E[X] = β. That is to say, the expected duration of survival of the system is β units of time. The parameterisation involving the "rate" parameter arises in the context of events arriving at a rate λ, when the time between events (which might be modelled using an exponential distribution) has a mean of β = λ−1.

The alternative specification is sometimes more convenient than the one given above, and some authors will use it as a standard definition. This alternative specification is not used here. Unfortunately this gives rise to a notational ambiguity. In general, the reader must check which of these two specifications is being used if an author writes "X ~ Exponential(λ)", since either the notation in the previous (using λ) or the notation in this section (here, using β to avoid confusion) could be intended.

## Properties

### Mean, variance, and median

The mean or expected value of an exponentially distributed random variable X with rate parameter λ is given by

$\mathrm{E}[X] = \frac{1}{\lambda}. \!$

In light of the examples given above, this makes sense: if you receive phone calls at an average rate of 2 per hour, then you can expect to wait half an hour for every call.

The variance of X is given by

$\mathrm{Var}[X] = \frac{1}{\lambda^2}. \!$

The median of X is given by

$\text{m}[X] = \frac{\ln 2}{\lambda} < \text{E}[X], \!$

where ln refers to the natural logarithm. Thus the absolute difference between the mean and median is

$| \text{E}[X]- \text{m}[X]| = \frac{1- \ln 2}{\lambda}< \frac{1}{\lambda} = \text{standard deviation},$

in accordance with the median-mean inequality.

### Memorylessness

An important property of the exponential distribution is that it is memoryless. This means that if a random variable T is exponentially distributed, its conditional probability obeys

$\Pr(T > s + t\; |\; T > s) = \Pr(T > t) \text{ for all } s, t \ge 0.$

This says that the conditional probability that we need to wait, for example, more than another 10 seconds before the first arrival, given that the first arrival has not yet happened after 30 seconds, is equal to the initial probability that we need to wait more than 10 seconds for the first arrival. So, if we waited for 30 seconds and the first arrival didn't happen (T > 30), probability that we'll need to wait another 10 seconds for the first arrival (T > 30 + 10) is the same as the initial probability that we need to wait more than 10 seconds for the first arrival (T > 10). Students taking courses on probability often misunderstand this idea: the fact that Pr(T > 40 | T > 30) = Pr(T > 10) does not mean that the events T > 40 and T > 30 are independent.

To summarize: "memorylessness" of the probability distribution of the waiting time T until the first arrival means

$\text{(Right) }\Pr(T>40 \mid T>30)=\Pr(T>10). \,$

It does not mean

$\text{(Wrong) }\Pr(T>40 \mid T>30)=\Pr(T>40). \,$

(That would be independence. These two events are not independent.)

The exponential distributions and the geometric distributions are the only memoryless probability distributions.

The exponential distribution is consequently also necessarily the only continuous probability distribution that has a constant Failure rate.

### Quantiles

The quantile function (inverse cumulative distribution function) for Exponential(λ) is

$F^{-1}(p;\lambda) = \frac{-\ln(1-p)}{\lambda}, \!$

for 0 ≤ p < 1. The quartiles are therefore:

first quartile
ln(4/3)/λ
median
ln(2)/λ
third quartile
ln(4)/λ

### Kullback–Leibler divergence

The directed Kullback–Leibler divergence between Exp(λ0) ('true' distribution) and Exp(λ) ('approximating' distribution) is given by

$\Delta(\lambda_0 || \lambda) = \log(\lambda_0) - \log(\lambda) + \frac{\lambda}{\lambda_0} - 1.$

### Maximum entropy distribution

Among all continuous probability distributions with support [0,∞) and mean μ, the exponential distribution with λ = 1/μ has the largest entropy. Alternatively, it is the maximum entropy probability distribution for a random variate X for which E(X) is fixed and greater than zero.[1]

### Distribution of the minimum of exponential random variables

Let X1, ..., Xn be independent exponentially distributed random variables with rate parameters λ1, ..., λn. Then

$\min\{\,X_1,\dots,X_n\,\}$

is also exponentially distributed, with parameter

$\lambda = \lambda_1+\cdots+\lambda_n.\,$

This can be seen by considering the complementary cumulative distribution function:

\begin{align} \Pr(\min\{\,X_1,\dots,X_n\,\} > x) & = \Pr\left(X_1 > x \text{ and }\dots\text{ and }X_n > x\right) \\ = \prod_{i=1}^n \Pr(X_i > x) & = \prod_{i=1}^n \exp(-x\lambda_i) = \exp\left(-x\sum_{i=1}^n \lambda_i\right). \end{align}

The index of the variable which achieves the minimum is distributed according to the law

$\Pr(X_k=\min\{\,X_1,\dots,X_n\,\})=\frac{\lambda_k}{\lambda_1+\cdots+\lambda_n}.$

Note that

$\max\{\,X_1,\dots,X_n\,\}$

is not exponentially distributed.

## Parameter estimation

Suppose a given variable is exponentially distributed and the rate parameter λ is to be estimated.

### Maximum likelihood

The likelihood function for λ, given an independent and identically distributed sample x = (x1, ..., xn) drawn from the variable, is

$L(\lambda) = \prod_{i=1}^n \lambda \, \exp(-\lambda x_i) = \lambda^n \, \exp\!\left(\!-\lambda \sum_{i=1}^n x_i\right)=\lambda^n\exp\left(-\lambda n \overline{x}\right),$

where

$\overline{x}={1 \over n}\sum_{i=1}^n x_i$

is the sample mean.

The derivative of the likelihood function's logarithm is

$\frac{\mathrm{d}}{\mathrm{d}\lambda} \ln L(\lambda) = \frac{\mathrm{d}}{\mathrm{d}\lambda} \left( n \ln(\lambda) - \lambda n\overline{x} \right) = {n \over \lambda}-n\overline{x}\ \begin{cases} > 0 & \text{if }0 < \lambda < 1/\overline{x}, \\[8pt] = 0 & \text{if }\lambda = 1/\overline{x}, \\[8pt] < 0 & \text{if }\lambda > 1/\overline{x}. \end{cases}$

Consequently the maximum likelihood estimate for the rate parameter is

$\widehat{\lambda} = \frac1{\overline{x}}.$

While this estimate is the most likely reconstruction of the true parameter λ, it is only an estimate, and as such, one can imagine that the more data points are available the better the estimate will be. It so happens that one can compute an exact confidence interval – that is, a confidence interval that is valid for all number of samples, not just large ones. The 100(1 − α)% exact confidence interval for this estimate is given by[2]

$\frac{2n}{\widehat{\lambda}\chi^2_{1-\alpha/2,2n}} < \frac{1}{\lambda} < \frac{2n}{\widehat{\lambda}\chi^2_{\alpha/2,2n}}$

which is also equal to:

$\frac{2n\overline{x}}{\chi^2_{1-\alpha/2,2n}} < \frac{1}{\lambda} < \frac{2n\overline{x}}{\chi^2_{\alpha/2,2n}}$

where $\widehat{\lambda}$ is the MLE estimate, λ is the true value of the parameter, and χ2p,ν is the 100(1 – p) percentile of the chi squared distribution with ν degrees of freedom.

### Bayesian inference

The conjugate prior for the exponential distribution is the gamma distribution (of which the exponential distribution is a special case). The following parameterization of the gamma pdf is useful:

$\mathrm{Gamma}(\lambda \,;\, \alpha, \beta) = \frac{\beta^{\alpha}}{\Gamma(\alpha)} \, \lambda^{\alpha-1} \, \exp(-\lambda\,\beta). \!$

The posterior distribution p can then be expressed in terms of the likelihood function defined above and a gamma prior:

\begin{align} & {} \qquad p(\lambda) \propto L(\lambda) \times \mathrm{Gamma}(\lambda \,;\, \alpha, \beta) \\ {} & = \lambda^n \, \exp(-\lambda\,n\overline{x}) \times \frac{\beta^{\alpha}}{\Gamma(\alpha)} \, \lambda^{\alpha-1} \, \exp(-\lambda\,\beta) \\ & {} \propto \lambda^{(\alpha+n)-1} \, \exp(-\lambda\,(\beta + n\overline{x})). \end{align}

Now the posterior density p has been specified up to a missing normalizing constant. Since it has the form of a gamma pdf, this can easily be filled in, and one obtains

$p(\lambda) = \mathrm{Gamma}(\lambda \,;\, \alpha + n, \beta + n \overline{x}).$

Here the parameter α can be interpreted as the number of prior observations, and β as the sum of the prior observations.

### Confidence interval

A simple and rapid method to calculate an approximate confidence interval for the estimation of λ is based on the application of the central limit theorem.[3] This method provides a good approximation of the confidence interval limits, for samples containing at least 15 – 20 elements. Denoting by N the sample size, the upper and lower limits of the 95% confidence interval are given by:

$\lambda_{low}=\widehat{\lambda} (1-1.96/\sqrt{N})$
$\lambda_{upp}=\widehat{\lambda} (1+1.96/\sqrt{N})$

## Generating exponential variates

A conceptually very simple method for generating exponential variates is based on inverse transform sampling: Given a random variate U drawn from the uniform distribution on the unit interval (0, 1), the variate

$T = F^{-1}(U) \!$

has an exponential distribution, where F −1 is the quantile function, defined by

$F^{-1}(p)=\frac{-\ln(1-p)}{\lambda}. \!$

Moreover, if U is uniform on (0, 1), then so is 1 − U. This means one can generate exponential variates as follows:

$T = \frac{-\ln U}{\lambda}. \!$

Other methods for generating exponential variates are discussed by Knuth[4] and Devroye.[5]

The ziggurat algorithm is a fast method for generating exponential variates.

A fast method for generating a set of ready-ordered exponential variates without using a sorting routine is also available.[5]

## Related distributions

• Exponential distribution is closed under scaling by a positive factor. If $X \sim \textrm{Exponential}(\lambda)\,$ then $kX \sim \textrm{Exponential}(\tfrac{\lambda}{k})\,$
• If $X \sim \textrm{Exponential}(\lambda_1)\,$ and $Y \sim \textrm{Exponential}(\lambda_2)\,$ then $\textrm{Min}(X,Y) \sim \textrm{Exponential}(\lambda_1+\lambda_2)\,$
• If $X_i \sim \textrm{Exponential}(\lambda)\,$ then $\textrm{Min}(X_{1,...,n}) \sim \textrm{Exponential}(n\lambda)\,$
• The Benktander Weibull distribution reduces to a truncated exponential distribution
• If $X \sim \textrm{Exponential}(\lambda)\,$ then $1+X \sim \textrm{BenktanderWeibull}(\lambda, 1)\,$ (Benktander Weibull distribution)
• The exponential distribution is a limit of a scaled beta distribution: $\lim_{n \to \infty}n{\rm Beta}(1,n) = {\rm Exponential}(1) \,$
• If $X_i \sim \mathrm{Exponential}(\lambda)\,$ then $\sum_{i=1}^k{X_i} \sim \textrm{Erlang}(k,\lambda)\,$ (Erlang distribution)
• If $X \sim \textrm{Exponential}(1)\,$ then $\mu - \sigma \log{X} \sim \textrm{Extreme}(\mu,\sigma)\,$ (Generalized extreme value distribution)
• If $X \sim \textrm{Exponential}(\lambda)\,$ then $X \sim \Gamma(1,\tfrac{1}{\lambda})\,$ (gamma distribution)
• If $X_1 \sim \mathrm{Exponential}(\lambda_1)\,$ and $X_2 \sim \mathrm{Exponential}(\lambda_2)\,$ then $\lambda_1 X_1-\lambda_2 X_2 \sim \mathrm{Laplace}\left(0,1\right)\,$ (Laplace distribution)
• If $X \sim \mathrm{Exponential}(\tfrac{1}{\lambda})\,$ and $Y \sim \mathrm{Exponential}(\tfrac{1}{\lambda})\,$ then $X-Y \sim \mathrm{Laplace}\left(0,\lambda\right)\,$
• If $X \sim \mathrm{Laplace}\left(\mu,\tfrac{1}{\beta}\right)\,$ then $|X-\mu| \sim \mathrm{Exponential}(\beta)\,$
• If $X \sim \mathrm{Exponential}(1)\,$ then $\mu-\beta\log{\tfrac{e^{-X}}{1-e^{-X}}} \sim \mathrm{Logistic}(\mu,\beta)$ (logistic distribution)
• If $X \sim \mathrm{Exponential}(1)\,$ and $Y \sim \mathrm{Exponential}(1)\,$ then $\mu-\beta\log{\tfrac{X}{Y}} \sim \mathrm{Logistic}(\mu,\beta)$ (logistic distribution)
• If $X \sim \mathrm{Exponential}(\alpha)\,$ then $k e^X \sim \mathrm{Pareto}(k,\alpha) \,$ (Pareto distribution)
• If $X \sim \mathrm{Pareto}(1,\lambda)\,$ then $\log{X} \sim \mathrm{Exponential}(\lambda) \,$
• Exponential distribution is a special case of type 3 Pearson distribution
• If $X \sim \mathrm{Exponential}(a)\,$ then $\tfrac{e^{-X}}{k} \sim \mathrm{PowerLaw}(k,a) \,$ (power law)
• If $X \sim \mathrm{Rayleigh}\left(\sqrt{\tfrac{1}{\lambda}}\right) \,$ then $\tfrac{1}{2}{X^2} \sim \mathrm{Exponential}(\lambda) \,$ (Rayleigh distribution)
• If $X \sim \mathrm{Exponential}(\lambda)\,$ then $X \sim \mathrm{Weibull}(\tfrac{1}{\lambda},1) \,$ (Weibull distribution)
• If $X \sim \mathrm{Exponential}(1)\,$ then $\lambda X^{\tfrac{1}{k}} \sim \mathrm{Weibull}(\lambda,k) \,$ (Weibull distribution)
• If $X_i \sim \mathrm{U}(0,1)\,$ (Uniform distribution (continuous)) then $\lim_{n \to \infty}n\textrm{Min}(X_{1,...,n}) \sim \textrm{Exponential}(1)\,$
• If $Y|X \sim \mathrm{Poisson}(X)\,$ (Poisson distribution) where $X \sim \mathrm{Exponential}(\tfrac{1}{\lambda})\,$ then $Y \sim \mathrm{Geometric}(\tfrac{1}{1+\lambda}) \,$ (geometric distribution)
• If $X \sim \mathrm{Exponential}(1)\,$ and $Y \sim \Gamma(\alpha,\tfrac{\beta}{\alpha}) \,$ then $\sqrt{XY} \sim \mathrm{K}(\alpha,\beta) \,$ (K-distribution)
• The Hoyt distribution can be obtained from Exponential distribution and Arcsine distribution
• If $X \sim \mathrm{Exponential}(\lambda)\,$ and $Y \sim \mathrm{Erlang}(n,\lambda)\,$ then $\tfrac{X}{Y} \sim \mathrm{Pareto}(1,n)\,$
• If $X \sim \mathrm{Exponential}(\lambda)\,$ and $Y \sim \Gamma(n,\tfrac{1}{\lambda}) \,$ then $\tfrac{X}{Y} \sim \mathrm{Pareto}(1,n)\,$
• Y ∼ Gumbel(μ, β), i.e. Y has a Gumbel distribution, if Y = μ − βlog() and X ∼ Exponential(λ).
• Xχ22, i.e. X has a chi-squared distribution with 2 degrees of freedom, if $X \sim \operatorname{Exponential}(\lambda = 1/2)\,\;$.
• X ∼ SkewLogistic(θ), then log(1 + e−−) ∼ Exponential(θ): see skew-logistic distribution.
• Let X ∼ Exponential(λX) and Y ∼ Exponential(λY) be independent. Then $Z = \frac{\lambda_X X}{\lambda_Y Y}$ has probability density function $f_Z(z) = \frac{1}{(z + 1)^2}$. This can be used to obtain a confidence interval for $\frac{\lambda_X}{\lambda_Y}$.

Other related distributions:

## Applications

### Occurrence of events

The exponential distribution occurs naturally when describing the lengths of the inter-arrival times in a homogeneous Poisson process.

The exponential distribution may be viewed as a continuous counterpart of the geometric distribution, which describes the number of Bernoulli trials necessary for a discrete process to change state. In contrast, the exponential distribution describes the time for a continuous process to change state.

In real-world scenarios, the assumption of a constant rate (or probability per unit time) is rarely satisfied. For example, the rate of incoming phone calls differs according to the time of day. But if we focus on a time interval during which the rate is roughly constant, such as from 2 to 4 p.m. during work days, the exponential distribution can be used as a good approximate model for the time until the next phone call arrives. Similar caveats apply to the following examples which yield approximately exponentially distributed variables:

• The time until a radioactive particle decays, or the time between clicks of a geiger counter
• The time it takes before your next telephone call
• The time until default (on payment to company debt holders) in reduced form credit risk modeling

Exponential variables can also be used to model situations where certain events occur with a constant probability per unit length, such as the distance between mutations on a DNA strand, or between roadkills on a given road.[citation needed]

In queuing theory, the service times of agents in a system (e.g. how long it takes for a bank teller etc. to serve a customer) are often modeled as exponentially distributed variables. (The inter-arrival of customers for instance in a system is typically modeled by the Poisson distribution in most management science textbooks.) The length of a process that can be thought of as a sequence of several independent tasks is better modeled by a variable following the Erlang distribution (which is the distribution of the sum of several independent exponentially distributed variables).

Fitted cumulative exponential distribution to annually maximum 1-day rainfalls using CumFreq [6]

Reliability theory and reliability engineering also make extensive use of the exponential distribution. Because of the memoryless property of this distribution, it is well-suited to model the constant hazard rate portion of the bathtub curve used in reliability theory. It is also very convenient because it is so easy to add failure rates in a reliability model. The exponential distribution is however not appropriate to model the overall lifetime of organisms or technical devices, because the "failure rates" here are not constant: more failures occur for very young and for very old systems.

In physics, if you observe a gas at a fixed temperature and pressure in a uniform gravitational field, the heights of the various molecules also follow an approximate exponential distribution. This is a consequence of the entropy property mentioned below.

In hydrology, the exponential distribution is used to analyze extreme values of such variables as monthly and annual maximum values of daily rainfall and river discharge volumes.[7]

The blue picture illustrates an example of fitting the exponential distribution to ranked annually maximum one-day rainfalls showing also the 90% confidence belt based on the binomial distribution. The rainfall data are represented by plotting positions as part of the cumulative frequency analysis.

### Prediction

Having observed a sample of n data points from an unknown exponential distribution a common task is to use these samples to make predictions about future data from the same source. A common predictive distribution over future samples is the so-called plug-in distribution, formed by plugging a suitable estimate for the rate parameter λ into the exponential density function. A common choice of estimate is the one provided by the principle of maximum likelihood, and using this yields the predictive density over a future sample xn+1, conditioned on the observed samples x = (x1, ..., xn) given by

$p_{\rm ML}(x_{n+1} | x_1, \ldots, x_n) = \left( \frac1{\overline{x}} \right) \exp \left( - \frac{x_{n+1}}{\overline{x}} \right)\!$

The Bayesian approach provides a predictive distribution which takes into account the uncertainty of the estimated parameter, although this may depend crucially on the choice of prior.

A predictive distribution free of the issues of choosing priors that arise under the subjective Bayesian approach is

$p_{\rm CNML}(x_{n+1} | x_1, \ldots, x_n) = \frac{ n^{n+1} \left( \overline{x} \right)^{n} }{ \left( n \overline{x} + x_{n+1} \right)^{n+1} }\!$,

which can be considered as (1) a frequentist confidence distribution, obtained from the distribution of the pivotal quantity ${x_{n+1}}/{\overline{x}}$;[8] (2) a profile predictive likelihood, obtained by eliminating the parameter λ from the joint likelihood of xn + 1 and λ by maximization;[9] (3) an objective Bayesian predictive posterior distribution, obtained using the non-informative Jeffreys prior 1 / λ; and (4) the Conditional Normalized Maximum Likelihood (CNML) predictive distribution, from information theoretic considerations.[10]

The accuracy of a predictive distribution may be measured using the distance or divergence between the true exponential distribution with rate parameter, λ0, and the predictive distribution based on the sample x. The Kullback–Leibler divergence is a commonly used, parameterisation free measure of the difference between two distributions. Letting Δ(λ0||p) denote the Kullback–Leibler divergence between an exponential with rate parameter λ0 and a predictive distribution p it can be shown that

\begin{align} {\rm E}_{\lambda_0} \left[ \Delta(\lambda_0||p_{\rm ML}) \right] &= \psi(n) + \frac{1}{n-1} - \log n \\ {\rm E}_{\lambda_0} \left[ \Delta(\lambda_0||p_{\rm CNML}) \right] &= \psi(n) + \frac{1}{n} - \log n \\ \end{align}

where the expectation is taken with respect to the exponential distribution with rate parameter λ0 ∈ (0, ∞), and ψ( · ) is the digamma function. It is clear that the CNML predictive distribution is strictly superior to the maximum likelihood plug-in distribution in terms of average Kullback–Leibler divergence for all sample sizes n > 0.

• Dead time – an application of exponential distribution to particle detector analysis.
• Laplace distribution, or the "double exponential distribution".

## References

1. ^ Park, Sung Y.; Bera, Anil K. (2009). "Maximum entropy autoregressive conditional heteroskedasticity model". Journal of Econometrics (Elsevier): 219–230. Retrieved 2011-06-02.
2. ^ K. S. Trivedi, Probability and Statistics with Reliability, Queueing and Computer Science applications, Chapter 10 Statistical Inference, http://www.ee.duke.edu/~kst/BLUEppt/chap10f_secure.pdf
3. ^ Guerriero V. et al. (2010). "Quantifying uncertainties in multi-scale studies of fractured reservoir analogues: Implemented statistical analysis of scan line data from carbonate rocks" (PDF). Journal of Structural Geology (Elsevier). doi:10.1016/j.jsg.2009.04.016.
4. ^ Donald E. Knuth (1998). The Art of Computer Programming, volume 2: Seminumerical Algorithms, 3rd edn. Boston: Addison–Wesley. ISBN 0-201-89684-2. See section 3.4.1, p. 133.
5. ^ a b Luc Devroye (1986). Non-Uniform Random Variate Generation. New York: Springer-Verlag. ISBN 0-387-96305-7. See chapter IX, section 2, pp. 392–401.
6. ^
7. ^ Ritzema (ed.), H.P. (1994). Frequency and Regression Analysis. Chapter 6 in: Drainage Principles and Applications, Publication 16, International Institute for Land Reclamation and Improvement (ILRI), Wageningen, The Netherlands. pp. 175–224. ISBN 90 70754 3 39.
8. ^ Lawless, J.F., Fredette, M.,"Frequentist predictions intervals and predictive distributions", Biometrika (2005), Vol 92, Issue 3, pp 529–542.
9. ^ Bjornstad, J.F., "Predictive Likelihood: A Review", Statist. Sci. Volume 5, Number 2 (1990), 242–254.
10. ^ D. F. Schmidt and E. Makalic, "Universal Models for the Exponential Distribution", IEEE Transactions on Information Theory, Volume 55, Number 7, pp. 3087–3090, 2009 doi:10.1109/TIT.2009.2018331

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