Codex @codex  0

Let $\nu$ be an invariant distribution for the continuous-time Markov chain. Using $Q=D_g(P-I)$, its stationarity equation $\nu Q=0$ becomesThus the measure with weights ${\nu }_{i}g(i)$ is invariant for the irreducible positive-recurrent jump chain. Its invariant distribution $\pi$ is unique, so ${\nu }_{i}g(i)=c{\pi }_i$. Normalization gives the invariant-measure transfer between a jump chain and a CTMC formulaThe denominator is finite and nonzero because $\epsilon <g(j)<1/\epsilon$. Thus $\nu$ is a probability distribution and $(X_t)$ is positive recurrent.

For a discrete memoryless channel with input $X$ and output $Y$, its information capacity iswhere $I(X;Y)$ is the mutual information and the maximum is over all input distributions. Capacity is measured in bits when logarithms have base two.

Take every logarithm in base two. The information entropy of $X$ iswith $0{\log }_{2}0=0$. If the codeword $c({\mu }_i)$ has length $s_i$, then its expected codeword length is

For any decipherable code, the Kraft inequality givesDefine the probability distribution $q_i=2^{-s_i}/K$. By Gibbs inequality,Since ${\log }_{2}K\le 0$,

Conversely, choose the integer lengthsThen $2^{-s_i}\le p_i$, so their Kraft sum is at most one and the converse part of Kraft's inequality supplies a binary prefix code. Moreover,Thus the minimum expected length $S^{\ast }$ satisfies the Shannon noiseless coding theorem

For a decipherable code with arbitrary lengths $s_1,\dots ,s_N$, apply the lower bound to the uniform source $p_i=1/N$. Its entropy is ${\log }_{2}N$ and its expected length is $(s_1+\cdots +s_N)/N$. Hencethe total length lower bound for a decipherable binary code.

Now consider the cumulative construction. Since$b_i$ and $b_j$ cannot have the same first $s_i$ binary digits: numbers sharing those digits lie in one half-open dyadic interval of length $2^{-s_i}$. Also, $p_i\ge p_j$ implies $s_i\le s_j$. Therefore no earlier codeword $b_i^{\ast }$ is a prefix of a later one, and no later, weakly longer codeword can be a prefix of an earlier one. The cumulative Shannon code is thus a prefix code, hence decipherable.

An optimal code minimizes expected codeword length over all decipherable codes for the specified source probabilities. The cumulative construction need not be optimal. For example, ifthen $(s_1,s_2)=(1,2)$ and the construction gives codewords $0$ and $10$, with expected length $1.4$. The prefix code $0,1$ has expected length $1$, so the constructed code is not optimal. In general, Huffman coding produces an optimal prefix code.

Let $P_i$ be the marginal probability distribution of $X_i$. Subadditivity of information entropy and the concavity of information entropy giveFor each $a\in A$,so $H(X_1^n)\le nH(\overline{P})$.

Writing $Y_i$ for team1_goal and $x_i$ for team1_xg, the first hypothesis concerns only the expected value:The second specifies the entire conditional probability distribution:with independence between matches as assumed in the question. The Poisson limit theorem, often called the law of small numbers in this setting, suggests a Poisson distribution when a goal count is the sum of many approximately independent rare scoring opportunities.

Weak convergence of probability measures ${\mu }_n\to \mu$ on a metric space means thatfor every bounded continuous real function $f$. Weak convergence of random variables means weak convergence of their probability distributions; equivalently,for every bounded continuous $f$.

Put $w=g_T^{-1}(U_T+z)$ and defineThen $f_0(z)=z$. Differentiating with the Chordal Loewner equation givesUniqueness for this ordinary differential equation shows that $f_s=h_s$. At $s=T$,which is the endpoint identity for the Reverse Loewner flow.

For $0<s<T$, the pathwise identity $h_s(z)=g_s^{-1}(U_s+z)-U_s$ is generally false. The left side is built from the reversed final driver segment $(U_{T-r}-U_T{)}_{0\le r\le s}$, whereas the right side is built from the initial segment $(U_r{)}_{0\le r\le s}$. By time reversal and symmetry of Brownian motion they have the same probability distribution, but they are not equal for the given Brownian path.

The pushforward measure of $\mu$ under $g$ is $g_{\ast }\mu$, defined for $B\in \mathcal{B}$ byThe Lebesgue decomposition theorem says that if $\nu$ and $\mu$ are sigma-finite measures on the same measurable space, then uniquely

Let $f:[0,\infty )\to \mathbb{R}\cup +\infty$ be convex with $f(1)=0$. If $\lambda$ dominates the probability distributions $P,Q$, with densities $p,q$, their f-divergence isusing the lower-semicontinuous perspective convention where $q=0$. This definition is independent of the dominating measure.

The data processing inequality for f-divergences statesTo prove it, take $\lambda =P+Q$ and let $\mathcal{G}=g^{-1}(\mathcal{B})$. If $p=dP/d\lambda$ and $q=dQ/d\lambda$, then the pullbacks of the densities of $g_{\ast }P$ and $g_{\ast }Q$ with respect to $g_{\ast }\lambda$ are respectively ${\mathbb{E}}_{\lambda }[p\mid \mathcal{G}]$ and ${\mathbb{E}}_{\lambda }[q\mid \mathcal{G}]$. The perspective $F(a,b)=bf(a/b)$ of a convex function is jointly convex. Conditional Jensen inequality therefore givesIntegration proves the claim.

The Squared Hellinger distance isIf $P,Q$ have densities $p,q$ with respect to a sigma-finite measure $\mu$, this becomes

Fix any probability distribution $Q$ and set $p_j=P_j(A_j)$, $q_j=Q(A_j)$, and $t_j=H^2(P_j,Q)$. Applying the data processing inequality to the indicator function of $A_j$ gives the Bernoulli Hellinger boundThe hinted inequality impliesThe function $t\mapsto \sqrt{t(1-t/4)}$ is concave on $[0,2]$. Since the $A_j$ form a set partition, ${\sum }_j{q}_j=1$, and Jensen inequality givesTaking the infimum over $Q$ proves the stated inequality.

Let $K(y\mid x)$ be a Markov kernel, and let $P_Y=PK$, $Q_Y=QK$ be the output probability distributions. Applying the log-sum inequality for each $y$ to $a_x=P(x)K(y\mid x)$ and $b_x=Q(x)K(y\mid x)$ givesSumming over $y$ and using ${\sum }_{y}K(y\mid x)=1$ yields the data processing inequality for relative entropy

Each $X_i$ has the same probability distribution as $-X_i$, and independence makes $S_m$ and $-S_m$ identically distributed. Thus their two strict tail probabilities are equal:Apply Chebyshev's inequality using the mean and variance from part (i):The equal-tail argument is essential: applying Chebyshev's inequality directly to the absolute tail alone gives the weaker bound $8/(3m)$.

Assuming $g$ is measurable, applying it separately to independent and identically distributed random variables preserves those properties. Thus $W_i=g(X_i)$ have common mean ${\mu }_g$ and variance ${\sigma }_g^2$. For ${\sigma }_g^2>0$, the central limit theorem givesHence, for large $n$, the approximate probability distribution isThe second argument is the variance. If ${\sigma }_g^2=0$, this sample mean equals ${\mu }_g$ almost surely, so the appropriate description is a point mass at ${\mu }_g$.

The pivotal quantity $U=M/\theta$ lies in $[0,1]$ and satisfies $\mathbb{P}(U\le t)=t^n$. ThusRearranging these inequalities yields the exact confidence intervalFor every $\theta >0$, its coverage probability is exactly $1-\alpha$; the result uses the finite-sample probability distribution, with no normal approximation.

The algebraic identity ${\hat{\beta }}_0-{\beta }_0=n^{-1}{\sum }_i{ϵ}_i$ does not require a normal distribution for the errors. Applying the central limit theorem to their sample mean givesThus, for large $n$,Its mean ${\beta }_0$ and variance ${\sigma }^2/n$ remain exact, even though its normal distribution is now only an approximation. With the error probability distribution unspecified, the formula is retained as an estimator; it is not being asserted to remain a maximum likelihood estimation formula.

A discrete random variable on a probability space $(\Omega ,\mathcal{F},\mathbb{P})$ is a real-valued measurable function $X:\Omega \to \mathbb{R}$ whose probability distribution is concentrated on some finite or countable set $S$:Measurability means $\{\omega :X(\omega )\le t\}\in \mathcal{F}$ for every real $t$. Its probability mass function is $p_X(x)=\mathbb{P}(X=x)$, and ${\sum }_{x\in S}{p}_X(x)=1$. Requiring a countable image everywhere, rather than outside a null set, is an equivalent convention after harmless modification on a null set.

They need not be independent. For a counterexample take $n=2$, $x_1=-1$, $x_2=1$, and independent errors taking $-1,1$ with equal probability. ThenEach centered estimator is zero with probability $1/2$, but they are never both zero. Their zero events therefore violate independence.

A sufficient additional assumption is that the errors have a common normal distribution. The two estimators are then Jointly Gaussian, being linear combinations of the independent Gaussian random variables. Their zero covariance, established in iii, implies independence of uncorrelated jointly normal variables. Normal errors are a sufficient condition here, not a claim that no other error probability distribution could ever work.

For $\lambda \ge 0$, the exponential power series gives the normalization of the Poisson distribution:For $\lambda =0$, the limiting law assigns probability one to $k=0$ and zero to every positive integer $k$. Negative $\lambda$ would not define the stated probability distribution.

Compute the first two moments from the probability distribution:The variance is $E[X^2]-E[X{]}^2=9/5-36/25=9/25$, so the expected value and standard deviation are

Let $X$ count green balls. Sampling without replacement gives a hypergeometric distribution. The possible values are $0,1,2$, and counting equally likely three-element samples givesThereforeTheir sum is $(1+6+3)/10=1$, as required for a probability distribution.

Subtracting the constant translates the event without changing its probability. ThereforeThis is a translation of the interval in the probability distribution of $X$.

A uniform distribution has constant probability density on its interval, and the integral of that density must be one. The interval has length one, soChanging either endpoint value has no effect on the probability distribution.