Exponential Families and Theoretical Inference
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Chapter 6 Continuous Random Variables and Probability
EF 507 QUANTITATIVE METHODS FOR ECONOMICS AND FINANCE FALL 2019 Chapter 6 Continuous Random Variables and Probability Distributions Chap 6-1 Probability Distributions Probability Distributions Ch. 5 Discrete Continuous Ch. 6 Probability Probability Distributions Distributions Binomial Uniform Hypergeometric Normal Poisson Exponential Chap 6-2/62 Continuous Probability Distributions § A continuous random variable is a variable that can assume any value in an interval § thickness of an item § time required to complete a task § temperature of a solution § height in inches § These can potentially take on any value, depending only on the ability to measure accurately. Chap 6-3/62 Cumulative Distribution Function § The cumulative distribution function, F(x), for a continuous random variable X expresses the probability that X does not exceed the value of x F(x) = P(X £ x) § Let a and b be two possible values of X, with a < b. The probability that X lies between a and b is P(a < X < b) = F(b) -F(a) Chap 6-4/62 Probability Density Function The probability density function, f(x), of random variable X has the following properties: 1. f(x) > 0 for all values of x 2. The area under the probability density function f(x) over all values of the random variable X is equal to 1.0 3. The probability that X lies between two values is the area under the density function graph between the two values 4. The cumulative density function F(x0) is the area under the probability density function f(x) from the minimum x value up to x0 x0 f(x ) = f(x)dx 0 ò xm where -
Skewed Double Exponential Distribution and Its Stochastic Rep- Resentation
EUROPEAN JOURNAL OF PURE AND APPLIED MATHEMATICS Vol. 2, No. 1, 2009, (1-20) ISSN 1307-5543 – www.ejpam.com Skewed Double Exponential Distribution and Its Stochastic Rep- resentation 12 2 2 Keshav Jagannathan , Arjun K. Gupta ∗, and Truc T. Nguyen 1 Coastal Carolina University Conway, South Carolina, U.S.A 2 Bowling Green State University Bowling Green, Ohio, U.S.A Abstract. Definitions of the skewed double exponential (SDE) distribution in terms of a mixture of double exponential distributions as well as in terms of a scaled product of a c.d.f. and a p.d.f. of double exponential random variable are proposed. Its basic properties are studied. Multi-parameter versions of the skewed double exponential distribution are also given. Characterization of the SDE family of distributions and stochastic representation of the SDE distribution are derived. AMS subject classifications: Primary 62E10, Secondary 62E15. Key words: Symmetric distributions, Skew distributions, Stochastic representation, Linear combina- tion of random variables, Characterizations, Skew Normal distribution. 1. Introduction The double exponential distribution was first published as Laplace’s first law of error in the year 1774 and stated that the frequency of an error could be expressed as an exponential function of the numerical magnitude of the error, disregarding sign. This distribution comes up as a model in many statistical problems. It is also considered in robustness studies, which suggests that it provides a model with different characteristics ∗Corresponding author. Email address: (A. Gupta) http://www.ejpam.com 1 c 2009 EJPAM All rights reserved. K. Jagannathan, A. Gupta, and T. Nguyen / Eur. -
1 One Parameter Exponential Families
1 One parameter exponential families The world of exponential families bridges the gap between the Gaussian family and general dis- tributions. Many properties of Gaussians carry through to exponential families in a fairly precise sense. • In the Gaussian world, there exact small sample distributional results (i.e. t, F , χ2). • In the exponential family world, there are approximate distributional results (i.e. deviance tests). • In the general setting, we can only appeal to asymptotics. A one-parameter exponential family, F is a one-parameter family of distributions of the form Pη(dx) = exp (η · t(x) − Λ(η)) P0(dx) for some probability measure P0. The parameter η is called the natural or canonical parameter and the function Λ is called the cumulant generating function, and is simply the normalization needed to make dPη fη(x) = (x) = exp (η · t(x) − Λ(η)) dP0 a proper probability density. The random variable t(X) is the sufficient statistic of the exponential family. Note that P0 does not have to be a distribution on R, but these are of course the simplest examples. 1.0.1 A first example: Gaussian with linear sufficient statistic Consider the standard normal distribution Z e−z2=2 P0(A) = p dz A 2π and let t(x) = x. Then, the exponential family is eη·x−x2=2 Pη(dx) / p 2π and we see that Λ(η) = η2=2: eta= np.linspace(-2,2,101) CGF= eta**2/2. plt.plot(eta, CGF) A= plt.gca() A.set_xlabel(r'$\eta$', size=20) A.set_ylabel(r'$\Lambda(\eta)$', size=20) f= plt.gcf() 1 Thus, the exponential family in this setting is the collection F = fN(η; 1) : η 2 Rg : d 1.0.2 Normal with quadratic sufficient statistic on R d As a second example, take P0 = N(0;Id×d), i.e. -
Parametric Models
An Introduction to Event History Analysis Oxford Spring School June 18-20, 2007 Day Two: Regression Models for Survival Data Parametric Models We’ll spend the morning introducing regression-like models for survival data, starting with fully parametric (distribution-based) models. These tend to be very widely used in social sciences, although they receive almost no use outside of that (e.g., in biostatistics). A General Framework (That Is, Some Math) Parametric models are continuous-time models, in that they assume a continuous parametric distribution for the probability of failure over time. A general parametric duration model takes as its starting point the hazard: f(t) h(t) = (1) S(t) As we discussed yesterday, the density is the probability of the event (at T ) occurring within some differentiable time-span: Pr(t ≤ T < t + ∆t) f(t) = lim . (2) ∆t→0 ∆t The survival function is equal to one minus the CDF of the density: S(t) = Pr(T ≥ t) Z t = 1 − f(t) dt 0 = 1 − F (t). (3) This means that another way of thinking of the hazard in continuous time is as a conditional limit: Pr(t ≤ T < t + ∆t|T ≥ t) h(t) = lim (4) ∆t→0 ∆t i.e., the conditional probability of the event as ∆t gets arbitrarily small. 1 A General Parametric Likelihood For a set of observations indexed by i, we can distinguish between those which are censored and those which aren’t... • Uncensored observations (Ci = 1) tell us both about the hazard of the event, and the survival of individuals prior to that event. -
6: the Exponential Family and Generalized Linear Models
10-708: Probabilistic Graphical Models 10-708, Spring 2014 6: The Exponential Family and Generalized Linear Models Lecturer: Eric P. Xing Scribes: Alnur Ali (lecture slides 1-23), Yipei Wang (slides 24-37) 1 The exponential family A distribution over a random variable X is in the exponential family if you can write it as P (X = x; η) = h(x) exp ηT T(x) − A(η) : Here, η is the vector of natural parameters, T is the vector of sufficient statistics, and A is the log partition function1 1.1 Examples Here are some examples of distributions that are in the exponential family. 1.1.1 Multivariate Gaussian Let X be 2 Rp. Then we have: 1 1 P (x; µ; Σ) = exp − (x − µ)T Σ−1(x − µ) (2π)p=2jΣj1=2 2 1 1 = exp − (tr xT Σ−1x + µT Σ−1µ − 2µT Σ−1x + ln jΣj) (2π)p=2 2 0 1 1 B 1 −1 T T −1 1 T −1 1 C = exp B− tr Σ xx +µ Σ x − µ Σ µ − ln jΣj)C ; (2π)p=2 @ 2 | {z } 2 2 A | {z } vec(Σ−1)T vec(xxT ) | {z } h(x) A(η) where vec(·) is the vectorization operator. 1 R T It's called this, since in order for P to normalize, we need exp(A(η)) to equal x h(x) exp(η T(x)) ) A(η) = R T ln x h(x) exp(η T(x)) , which is the log of the usual normalizer, which is the partition function. -
The Probability Lifesaver: Order Statistics and the Median Theorem
The Probability Lifesaver: Order Statistics and the Median Theorem Steven J. Miller December 30, 2015 Contents 1 Order Statistics and the Median Theorem 3 1.1 Definition of the Median 5 1.2 Order Statistics 10 1.3 Examples of Order Statistics 15 1.4 TheSampleDistributionoftheMedian 17 1.5 TechnicalboundsforproofofMedianTheorem 20 1.6 TheMedianofNormalRandomVariables 22 2 • Greetings again! In this supplemental chapter we develop the theory of order statistics in order to prove The Median Theorem. This is a beautiful result in its own, but also extremely important as a substitute for the Central Limit Theorem, and allows us to say non- trivial things when the CLT is unavailable. Chapter 1 Order Statistics and the Median Theorem The Central Limit Theorem is one of the gems of probability. It’s easy to use and its hypotheses are satisfied in a wealth of problems. Many courses build towards a proof of this beautiful and powerful result, as it truly is ‘central’ to the entire subject. Not to detract from the majesty of this wonderful result, however, what happens in those instances where it’s unavailable? For example, one of the key assumptions that must be met is that our random variables need to have finite higher moments, or at the very least a finite variance. What if we were to consider sums of Cauchy random variables? Is there anything we can say? This is not just a question of theoretical interest, of mathematicians generalizing for the sake of generalization. The following example from economics highlights why this chapter is more than just of theoretical interest. -
Lecture 2 — September 24 2.1 Recap 2.2 Exponential Families
STATS 300A: Theory of Statistics Fall 2015 Lecture 2 | September 24 Lecturer: Lester Mackey Scribe: Stephen Bates and Andy Tsao 2.1 Recap Last time, we set out on a quest to develop optimal inference procedures and, along the way, encountered an important pair of assertions: not all data is relevant, and irrelevant data can only increase risk and hence impair performance. This led us to introduce a notion of lossless data compression (sufficiency): T is sufficient for P with X ∼ Pθ 2 P if X j T (X) is independent of θ. How far can we take this idea? At what point does compression impair performance? These are questions of optimal data reduction. While we will develop general answers to these questions in this lecture and the next, we can often say much more in the context of specific modeling choices. With this in mind, let's consider an especially important class of models known as the exponential family models. 2.2 Exponential Families Definition 1. The model fPθ : θ 2 Ωg forms an s-dimensional exponential family if each Pθ has density of the form: s ! X p(x; θ) = exp ηi(θ)Ti(x) − B(θ) h(x) i=1 • ηi(θ) 2 R are called the natural parameters. • Ti(x) 2 R are its sufficient statistics, which follows from NFFC. • B(θ) is the log-partition function because it is the logarithm of a normalization factor: s ! ! Z X B(θ) = log exp ηi(θ)Ti(x) h(x)dµ(x) 2 R i=1 • h(x) 2 R: base measure. -
Statistical Inference
GU4204: Statistical Inference Bodhisattva Sen Columbia University February 27, 2020 Contents 1 Introduction5 1.1 Statistical Inference: Motivation.....................5 1.2 Recap: Some results from probability..................5 1.3 Back to Example 1.1...........................8 1.4 Delta method...............................8 1.5 Back to Example 1.1........................... 10 2 Statistical Inference: Estimation 11 2.1 Statistical model............................. 11 2.2 Method of Moments estimators..................... 13 3 Method of Maximum Likelihood 16 3.1 Properties of MLEs............................ 20 3.1.1 Invariance............................. 20 3.1.2 Consistency............................ 21 3.2 Computational methods for approximating MLEs........... 21 3.2.1 Newton's Method......................... 21 3.2.2 The EM Algorithm........................ 22 1 4 Principles of estimation 23 4.1 Mean squared error............................ 24 4.2 Comparing estimators.......................... 25 4.3 Unbiased estimators........................... 26 4.4 Sufficient Statistics............................ 28 5 Bayesian paradigm 33 5.1 Prior distribution............................. 33 5.2 Posterior distribution........................... 34 5.3 Bayes Estimators............................. 36 5.4 Sampling from a normal distribution.................. 37 6 The sampling distribution of a statistic 39 6.1 The gamma and the χ2 distributions.................. 39 6.1.1 The gamma distribution..................... 39 6.1.2 The Chi-squared distribution.................. 41 6.2 Sampling from a normal population................... 42 6.3 The t-distribution............................. 45 7 Confidence intervals 46 8 The (Cramer-Rao) Information Inequality 51 9 Large Sample Properties of the MLE 57 10 Hypothesis Testing 61 10.1 Principles of Hypothesis Testing..................... 61 10.2 Critical regions and test statistics.................... 62 10.3 Power function and types of error.................... 64 10.4 Significance level............................ -
Likelihood-Based Inference
The score statistic Inference Exponential distribution example Likelihood-based inference Patrick Breheny September 29 Patrick Breheny Survival Data Analysis (BIOS 7210) 1/32 The score statistic Univariate Inference Multivariate Exponential distribution example Introduction In previous lectures, we constructed and plotted likelihoods and used them informally to comment on likely values of parameters Our goal for today is to make this more rigorous, in terms of quantifying coverage and type I error rates for various likelihood-based approaches to inference With the exception of extremely simple cases such as the two-sample exponential model, exact derivation of these quantities is typically unattainable for survival models, and we must rely on asymptotic likelihood arguments Patrick Breheny Survival Data Analysis (BIOS 7210) 2/32 The score statistic Univariate Inference Multivariate Exponential distribution example The score statistic Likelihoods are typically easier to work with on the log scale (where products become sums); furthermore, since it is only relative comparisons that matter with likelihoods, it is more meaningful to work with derivatives than the likelihood itself Thus, we often work with the derivative of the log-likelihood, which is known as the score, and often denoted U: d U (θ) = `(θjX) X dθ Note that U is a random variable, as it depends on X U is a function of θ For independent observations, the score of the entire sample is the sum of the scores for the individual observations: X U = Ui i Patrick Breheny Survival -
Robust Estimation on a Parametric Model Via Testing 3 Estimators
Bernoulli 22(3), 2016, 1617–1670 DOI: 10.3150/15-BEJ706 Robust estimation on a parametric model via testing MATHIEU SART 1Universit´ede Lyon, Universit´eJean Monnet, CNRS UMR 5208 and Institut Camille Jordan, Maison de l’Universit´e, 10 rue Tr´efilerie, CS 82301, 42023 Saint-Etienne Cedex 2, France. E-mail: [email protected] We are interested in the problem of robust parametric estimation of a density from n i.i.d. observations. By using a practice-oriented procedure based on robust tests, we build an estimator for which we establish non-asymptotic risk bounds with respect to the Hellinger distance under mild assumptions on the parametric model. We show that the estimator is robust even for models for which the maximum likelihood method is bound to fail. A numerical simulation illustrates its robustness properties. When the model is true and regular enough, we prove that the estimator is very close to the maximum likelihood one, at least when the number of observations n is large. In particular, it inherits its efficiency. Simulations show that these two estimators are almost equal with large probability, even for small values of n when the model is regular enough and contains the true density. Keywords: parametric estimation; robust estimation; robust tests; T-estimator 1. Introduction We consider n independent and identically distributed random variables X1,...,Xn de- fined on an abstract probability space (Ω, , P) with values in the measure space (X, ,µ). E F We suppose that the distribution of Xi admits a density s with respect to µ and aim at estimating s by using a parametric approach. -
Chapter 10 “Some Continuous Distributions”.Pdf
CHAPTER 10 Some continuous distributions 10.1. Examples of continuous random variables We look at some other continuous random variables besides normals. Uniform distribution A continuous random variable has uniform distribution if its density is f(x) = 1/(b a) − if a 6 x 6 b and 0 otherwise. For a random variable X with uniform distribution its expectation is 1 b a + b EX = x dx = . b a ˆ 2 − a Exponential distribution A continuous random variable has exponential distribution with parameter λ > 0 if its λx density is f(x) = λe− if x > 0 and 0 otherwise. Suppose X is a random variable with an exponential distribution with parameter λ. Then we have ∞ λx λa (10.1.1) P(X > a) = λe− dx = e− , ˆa λa FX (a) = 1 P(X > a) = 1 e− , − − and we can use integration by parts to see that EX = 1/λ, Var X = 1/λ2. Examples where an exponential random variable is a good model is the length of a telephone call, the length of time before someone arrives at a bank, the length of time before a light bulb burns out. Exponentials are memoryless, that is, P(X > s + t X > t) = P(X > s), | or given that the light bulb has burned 5 hours, the probability it will burn 2 more hours is the same as the probability a new light bulb will burn 2 hours. Here is how we can prove this 129 130 10. SOME CONTINUOUS DISTRIBUTIONS P(X > s + t) P(X > s + t X > t) = | P(X > t) λ(s+t) e− λs − = λt = e e− = P(X > s), where we used Equation (10.1.1) for a = t and a = s + t. -
A PARTIALLY PARAMETRIC MODEL 1. Introduction a Notable
A PARTIALLY PARAMETRIC MODEL DANIEL J. HENDERSON AND CHRISTOPHER F. PARMETER Abstract. In this paper we propose a model which includes both a known (potentially) nonlinear parametric component and an unknown nonparametric component. This approach is feasible given that we estimate the finite sample parameter vector and the bandwidths simultaneously. We show that our objective function is asymptotically equivalent to the individual objective criteria for the parametric parameter vector and the nonparametric function. In the special case where the parametric component is linear in parameters, our single-step method is asymptotically equivalent to the two-step partially linear model esti- mator in Robinson (1988). Monte Carlo simulations support the asymptotic developments and show impressive finite sample performance. We apply our method to the case of a partially constant elasticity of substitution production function for an unbalanced sample of 134 countries from 1955-2011 and find that the parametric parameters are relatively stable for different nonparametric control variables in the full sample. However, we find substan- tial parameter heterogeneity between developed and developing countries which results in important differences when estimating the elasticity of substitution. 1. Introduction A notable development in applied economic research over the last twenty years is the use of the partially linear regression model (Robinson 1988) to study a variety of phenomena. The enthusiasm for the partially linear model (PLM) is not confined to