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Consider two binary random variables with local characteristics $$ \begin{aligned} &\operatorname{Pr}\left(Y_{1}=1 \mid Y_{2}=0\right)=\operatorname{Pr}\left(Y_{1}=0 \mid Y_{2}=1\right)=1 \\ &\operatorname{Pr}\left(Y_{2}=0 \mid Y_{1}=0\right)=\operatorname{Pr}\left(Y_{2}=1 \mid Y_{1}=1\right)=1 \end{aligned} $$ Show that these do not determine a joint density for \(\left(Y_{1}, Y_{2}\right) .\) Is the positivity condition satisfied?

Short Answer

Expert verified
A valid joint distribution isn't possible; the positivity condition is not satisfied due to contradictions.

Step by step solution

01

Understanding the Given Probabilities

We are given that: \( \operatorname{Pr}(Y_1=1 \mid Y_2=0) = 1 \) and \( \operatorname{Pr}(Y_1=0 \mid Y_2=1) = 1 \). These imply that if \( Y_2=0 \), then \( Y_1 \) is surely 1, and if \( Y_2=1 \), then \( Y_1 \) is surely 0.Additionally, \( \operatorname{Pr}(Y_2=0 \mid Y_1=0) = 1 \) and \( \operatorname{Pr}(Y_2=1 \mid Y_1=1) = 1 \). This means that if \( Y_1=0 \), \( Y_2 \) is sure to be 0, and if \( Y_1=1 \), \( Y_2 \) is sure to be 1.
02

Inferring Conditional Relationships

From the given probabilities, we can infer a relationship:- When \( Y_2=0 \), \( Y_1 \) must be 1.- When \( Y_2=1 \), \( Y_1 \) must be 0.- When \( Y_1=0 \), \( Y_2 \) must be 0.- When \( Y_1=1 \), \( Y_2 \) must be 1.These imply a deterministic relationship where \( Y_2 = 1 - Y_1 \), suggesting interdependence where exactly one of them needs to be 0 and the other 1.
03

Exploring Joint Distribution Possibility

A joint density function \( \operatorname{Pr}(Y_1, Y_2) \) for binary variables \( Y_1 \) and \( Y_2 \) must satisfy all marginal and conditional properties given by the problem. However, there seems to be an inconsistency:1. If \( Y_1 = 1 \), it implies \( Y_2 = 1 \).2. If \( Y_2 = 1 \), it implies \( Y_1 = 0 \).This reveals a contradiction: \( Y_1 \) can't be both 1 and 0 simultaneously when \( Y_2 = 1 \), so a valid joint distribution isn't possible due to overlapping constraints.
04

Checking Positivity Condition

The positivity condition for a valid probability distribution requires that all joint probabilities \( \operatorname{Pr}(Y_1=i, Y_2=j) \) be non-negative. In this scenario, establishing non-negative probabilities isn't possible for all required outcomes because of the contradictions between conditions derived. Thus, the positivity condition is not satisfied by these local characteristics.

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Key Concepts

These are the key concepts you need to understand to accurately answer the question.

Conditional Probability
Conditional probability is a crucial concept in probability theory, allowing us to calculate the likelihood of an event occurring given that another event has already occurred. It's generally denoted as \( \operatorname{Pr}(A \mid B) \), which represents the probability of event \( A \) happening given that \( B \) has already happened.
In the exercise, we deal with binary random variables \( Y_1 \) and \( Y_2 \). Being binary variables means each variable can take one of two possible outcomes, typically 0 or 1. The problem gives us several conditional probabilities:
  • \( \operatorname{Pr}(Y_1=1 \mid Y_2=0) = 1 \): If \( Y_2 \) is 0, \( Y_1 \) is 1 with certainty.
  • \( \operatorname{Pr}(Y_1=0 \mid Y_2=1) = 1 \): If \( Y_2 \) is 1, \( Y_1 \) must be 0.
  • \( \operatorname{Pr}(Y_2=0 \mid Y_1=0) = 1 \): If \( Y_1 \) is 0, \( Y_2 \) must also be 0.
  • \( \operatorname{Pr}(Y_2=1 \mid Y_1=1) = 1 \): If \( Y_1 \) is 1, \( Y_2 \) must be 1.
These conditional probabilities suggest that whenever one variable is known, the other is determined completely, showcasing a deterministic relationship, rather than a stochastic one. Understanding these conditional statements helps to identify whether a joint probability distribution can be established.
Binary Random Variables
Binary random variables are a specific type of random variable that take on one of exactly two possible values. These values are often labeled as 0 and 1. This binary nature makes them simple yet powerful tools for modeling various phenomena.
The exercise involves two binary random variables, \( Y_1 \) and \( Y_2 \). The conditional probability relationships given in the problem suggest that these variables have a perfectly deterministic relation: when one is known, the other's state is entirely predictable.
For example:
  • If you know \( Y_2=0 \), then \( Y_1 \) must be 1.
  • If \( Y_2=1 \), then \( Y_1 \) has to be 0.
These deterministic outcomes signify that any randomness typically expected in joint probability distributions is absent. Instead, it highlights a situation where one variable dictates the value of the other fully. This is contrary to what we usually expect with random variables, where there's some inherent uncertainty involved.
Positivity Condition
The positivity condition is an essential requirement for forming a valid probability distribution. It demands that all probabilities involved in the distribution must be non-negative and, preferably, positive.
In the provided scenario, we attempted to construct a joint probability distribution based on the given local characteristics. However, the deterministic nature of these relationships led to contradictions. A joint probability distribution must normally allow for simultaneous events (like \( Y_1=1 \) and \( Y_2=1 \)), but here we see contradictions. The variables suggest:
  • \( Y_1=1 \) implies \( Y_2=1 \)
  • \( Y_2=1 \) implies \( Y_1=0 \)
These conditions cannot occur together, which indicates inconsistency in forming valid joint probabilities. Since we cannot assign a valid positive joint probability to every possible outcome without contradicting another, the positivity condition fails here.
Essentially, the contradictions mean that valid probabilities can't be established for all conditions required, making the joint distribution impossible. Understanding this helps underscore why such strictitude is applied in forming joint probability densities, ensuring they reflect realistic probabilistic situations.

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Most popular questions from this chapter

One way to estimate the evolutionary distance between species is to identify sections of their DNA which are similar and so must derive from a common ancestor species. If such sections differ at very few sites, the species are closely related and must have separated recently in the evolutionary past, but if the sections differ by more, the species are further apart. For example, data from the first introns of human and owl monkey insulin genes are in Table \(6.12\). The first row means that there are 20 sites with \(\mathrm{A}\) on both genes, 0 with A on the human and C on the monkey, and so on. If all the data lay on the diagonal, this section would be identical in both species. Note that even if sites on both genes have the same base, there could have been changes such as (ancestor) \(\mathrm{A} \rightarrow \mathrm{G} \rightarrow \mathrm{T}\) (human) and (ancestor) \(\mathrm{A} \rightarrow \mathrm{C} \rightarrow \mathrm{A} \rightarrow \mathrm{T}\) (monkey). Here is a (greatly simplified) model for evolutionary distance. We suppose that at a time \(t_{0}\) in the past the two species we now see began to evolve away from a common ancestor species, which had a section of DNA of length \(n\) similar to those we now see. Each site on that section had one of the four bases \(\mathrm{A}, \mathrm{C}, \mathrm{G}\), or \(\mathrm{T}\), and for each species the base at each site has since changed according to a continuous-time Markov chain with infinitesimal generator $$ G=\left(\begin{array}{cccc} -3 \gamma & \gamma & \gamma & \gamma \\ \gamma & -3 \gamma & \gamma & \gamma \\ \gamma & \gamma & -3 \gamma & \gamma \\ \gamma & \gamma & \gamma & -3 \gamma \end{array}\right) $$ independent of other sites. That is, the rate at which one base changes into, or is substituted by, another is the same for any pair of bases. (a) Check that \(G\) has eigendecomposition $$ \frac{1}{4}\left(\begin{array}{cccc} 1 & -1 & -1 & -1 \\ 1 & -1 & -1 & 3 \\ 1 & -1 & 3 & -1 \\ 1 & 3 & -1 & -1 \end{array}\right)\left(\begin{array}{cccc} 0 & 0 & 0 & 0 \\ 0 & -4 \gamma & 0 & 0 \\ 0 & 0 & -4 \gamma & 0 \\ 0 & 0 & 0 & -4 \gamma \end{array}\right)\left(\begin{array}{cccc} 1 & 1 & 1 & 1 \\ -1 & 0 & 0 & 1 \\ -1 & 0 & 1 & 0 \\ -1 & 1 & 0 & 0 \end{array}\right) $$ find its equilibrium distribution \(\pi\), and show that the chain is reversible. (b) Show that exp \(t G)\) has diagonal elements \(\left(1+3 e^{-4 \gamma t}\right) / 4\) and off-diagonal elements \(\left(1-e^{-4 \gamma t}\right) / 4\). Use this and reversibility of the chain to explain why the likelihood for \(\gamma\) based on data like those above is proportional to $$ \left(1+3 e^{-8 \gamma t_{0}}\right)^{n-R}\left(1-e^{-8 \gamma t_{0}}\right)^{R} $$ where \(R\) is the number of sites at which the two sections disagree. Hence find an estimate and standard error for \(\gamma t_{0}\) for the data above. (c) Show that for each site, the probability of no substitution on either species in period \(t\) is \(1-\exp (-6 \gamma t)\), deduce that substitutions occur as a Poisson process of rate \(6 \gamma\), and hence show that the estimated mean number of substitutions per site for the data above is \(0.120\) Discuss the fit of this model.

Consider a Poisson process of intensity \(\lambda\) in the plane. Find the distribution of the area of the largest disk centred on one point but containing no other points.

Dataframe alof i contains three-state data derived from daily rainfall over three years at Alofi in the Niue Island group in the Pacific Ocean. The states are 1 (no rain), 2 (up to \(5 \mathrm{~mm}\) rain) and 3 (over \(5 \mathrm{~mm}\) ). Triplets of transition counts for all 1096 observations are given in the upper part of Table \(6.10 ;\) its lower part gives transition counts for successive pairs for sub-sequences 1-274, 275-548, 549-822 and 823-1096. (a) The maximized log likelihoods for first-, second-, and third-order Markov chains fitted to the entire dataset are \(-1038.06,-1025.10\), and \(-1005.56\). Compute the log likelihood for the zeroth-order model, and compare the four fits using likelihood ratio statistics and using AIC. Give the maximum likelihood estimates for the best-fitting model. Does it simplify to a varying-order chain? (b) Matrices of transition counts \(\left\\{n_{i r s}\right\\}\) are available for \(m\) independent \(S\)-state chains with transition matrices \(P_{i}=\left(p_{i r s}\right), i=1, \ldots, m .\) Show that the maximum likelihood estimates are \(\widehat{p}_{i r s}=n_{\text {irs }} / n_{i \cdot s}\), where \(\cdot\) denotes summation over the corresponding index. Show that the maximum likelihood estimates under the simpler model in which \(P_{1}=\cdots=P_{m}=\left(p_{r s}\right)\) are \(\widehat{p}_{r s}=n_{\cdot r s} / n_{\cdot s} .\) Deduce that the likelihood ratio statistic to compare these models is \(2 \sum_{i, r, s} n_{\text {irs }} \log \left(\widehat{p}_{i r s} / \widehat{p}_{r s}\right)\) and give its degrees of freedom. (c) Consider the lower part of Table 6.10. Explain how to use the statistic from (b) to test for equal transition probabilities in each section, and hence check stationarity of the data.

Let \(Y_{(1)}<\cdots0, \lambda>0 .\) Show that for \(r=2, \ldots, n\) $$ \operatorname{Pr}\left(Y_{(r)}>y \mid Y_{(1)}, \ldots, Y_{(r-1)}\right)=\exp \left\\{-\lambda r\left(y-y_{(r-1)}\right)\right\\}, \quad y>y_{(r-1)} $$ and deduce that the order statistics from a general continuous distribution form a Markov process.

Over the centuries natural disasters in a particular country have occurred as a Poisson process of rate \(\lambda(t)\). Any disaster at time \(t\) is known to have occurred only with probability \(\pi(t)\), due to the patchiness of historical records. If records of different disasters are preserved independently, show that the point process of known disasters is Poisson with intensity \(\lambda(t) \pi(t)\)

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