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Data augmentation enlarges a posterior distribution with latent variables whose removal by marginalization recovers the original posterior distribution. Convenient full conditional distributions then permit a Gibbs sampler on the enlarged space.

A single-coordinate Gibbs sampler update preserves the remaining coordinates and independently redraws the selected coordinate from its full conditional distribution. The joint measure of old and new states factors into a marginal and two identical conditional factors, making it symmetric. A fixed-probability mixture of such kernels therefore satisfies detailed balance.

If a model configuration leaves a parameter absent from its likelihood function, giving that parameter an independent flat improper prior makes the posterior distribution normalization infinite whenever the configuration has positive prior distribution mass. Integrate first over that parameter: a positive constant likelihood function factor is repeated over all of $\mathbb{R}$. This occurs for the second segment mean of a Gaussian change-point detection model when the proposed split is after the final observation. There is no posterior distribution or valid Gibbs sampler for that joint kernel. A proper prior distribution on the unused parameter is necessary if the no-change configuration is retained.

A full conditional distribution is the conditional law of one component of a joint model given all the other components and the observed data. A Gibbs sampler updates components by drawing from these laws.

For independent means $\theta \sim N({\mu }_{\theta },t_{\theta })$ and $\varphi \sim N({\mu }_{\varphi },t_{\varphi })$, the first mean's conditional precision is $m/v+1/t_{\theta }$ and conditional precision-weighted mean is ${\sum }_{i\le m}{x}_i/v+{\mu }_{\theta }/t_{\theta }$; replace $m$ by $n-m$ for the second mean. With $v\sim \mathrm{IG}(a_0,b_0)$ its conditional law is $\mathrm{IG}(a_0+n/2,b_0+S_m/2)$. The discrete split has conditional probabilities proportional to $e^{-S_m/(2v)}$. These full conditional distributions give a Gibbs sampler including $m=n$. Averaging the sampled indicators of that configuration estimates its probability; averaging its conditional probabilities gives an estimate by Rao-Blackwellization.

An observed astronomical colour is the sum of a normal intrinsic colour, a nonnegative dust contribution with an exponential distribution, and normal measurement noise. With latent $C_s\sim N(\mu ,v)$, $E_s\sim \mathrm{Exp}(\text{mean }\tau )$ and $O_s\mid C_s,E_s\sim N(C_s+E_s,r_s)$, the Gibbs sampler conditionals for $C_s$ are normal and those for $E_s$ are truncated normal distributions. A flat prior on $\mu$ and inverse-gamma priors on $v,\tau$ give inverse-gamma conditional updates for the two scales. Log-flat priors on both scales instead make the joint posterior improper when all $r_s>0$, despite formally proper full conditionals at generic latent states: the marginal observed likelihood has a positive limit at either zero-scale boundary. Proper positive-scale priors restore a genuine joint posterior.

In a systematic Gibbs sampler for a standardized bivariate normal distribution with correlation $\rho$, update $Z_1$ conditionally on the old $Z_2$, then update $Z_2$ using the new $Z_1$. Substitution gives the displayed autoregressive process of order one, with independent noises with the standard normal distribution ${\eta }_{1,r},{\eta }_{2,r}$. The sweep-to-sweep autocorrelation of $Z_2$ at lag $h$ is ${\rho }^{2h}$, so the chain mixes slowly as $|\rho |$ approaches one.

The full conditional distributions are $\lambda \mid m,x\sim \mathrm{Gamma}(\alpha +S_m,\beta +m)$ and $\varphi \mid m,x\sim \mathrm{Gamma}(\gamma +S_N-S_m,\delta +N-m)$, in shape-rate notation. Conditional on the two rates, $m$ has weights $w_j=\exp [j(\varphi -\lambda )+S_j\log (\lambda /\varphi )]$. A Gibbs sampler cycles through these three draws. Each update preserves the joint posterior distribution; subtracting the largest log weight before exponentiation stabilizes the categorical draw. Averages of $1_{\{m=N\}}$ estimate the posterior probability of no observed change.

A random-effect model with positive marginal likelihood at scale zero cannot yield a proper posterior distribution under a prior proportional to the reciprocal scale near zero. On a compact set of the other parameters, continuity and positivity give a positive likelihood lower bound, while ${\int }_0^{\epsilon }d\lambda /\lambda =\infty$. Individual full conditional distributions can still be proper, so their existence alone does not justify a Gibbs sampler with a probability distribution as its joint target.

With independent prior distributions $\beta \sim N(\mu ,\Sigma )$ and $s\sim \mathrm{IG}(a,b)$ in a normal distribution regression model, the conditional posterior distribution of $\beta$ has covariance matrix $V=(X^{T}X/s+{\Sigma }^{-1}{)}^{-1}$ and mean $V(X^{T}y/s+{\Sigma }^{-1}\mu )$. The conditional posterior distribution of $s$ is inverse-gamma distribution $\mathrm{IG}(a+n/2,b+\Vert y-X\beta {\Vert }^2/2)$. The shape contains no half-dimension contribution from $\beta$, since its prior distribution is independent of $s$. These conditionals give a two-block Gibbs sampler.

If the positive baseline rate has a gamma distribution of shape $a$ and rate $b$, its logarithm has this probability density function with respect to $d\mu$. The extra factor $e^{\mu }$ is the Jacobian determinant of the transformation. After a Gibbs sampler draws the conditional baseline rate, taking its logarithm is an exact update of the intercept. Omitting the Jacobian changes the prior.

In a normal linear model $Y\mid b,s\sim N(Xb,sI)$, assign the coefficient vector $b$ a proper normal distribution with fixed positive-definite covariance matrix $D$, independently of the variance $s$, whose prior distribution is inverse-gamma distribution. Completing the square givesMultiplying the residual likelihood function by the inverse-gamma distribution density also givesThese full conditional distributions permit a Gibbs sampler. The marginal coefficient posterior distribution is obtained by marginalization over $s$. Its conditional mean and covariance matrix both depend on $s$, so it is generally not a Student's t-distribution. Even if the columns of $X$ and the prior coordinates make the conditional coefficient components independent, averaging over a shared unknown $s$ need not preserve that independence.

A normal-inverse-gamma prior instead scales the conditional coefficient prior covariance matrix by $s$, producing a multivariate Student's t-distribution on marginalization. It is a different joint prior distribution.

For the Metropolis–Hastings algorithm, start at a point $x$ with positive target probability density function. Draw $Y$ from a proposal probability density function $q(y\mid x)$, generate an independent uniform distribution value $U$, and set the next state to $Y$ ifOtherwise retain $x$, including that repeated state in the sample. Where a proposed forward move is possible but its reverse probability density function vanishes, its acceptance probability is zero. Unknown normalizing constants in $\pi$ cancel.

For distinct states, the accepted probability flow iswhich is symmetric in $x,y$. This detailed balance identity, together with the holding probability, proves that $\pi$ is invariant. Under appropriate irreducibility and aperiodicity conditions the chain converges to $\pi$; successive states are generally dependent.

The Gibbs sampler updates coordinates from their full conditional distributions. In a systematic sweep, draw successivelyEach update preserves the target joint law: integrating the old coordinate against its conditional probability density function and replacing it by an independent draw from that same conditional leaves the joint probability density function unchanged. The composition of the coordinate updates therefore also preserves $\pi$. A Gibbs sampler coordinate move can be regarded as a Metropolis–Hastings algorithm move with acceptance probability one. Systematic sweeps preserve stationarity but need not themselves be reversible. Initialization away from stationarity requires convergence before treating the draws as approximately target-distributed, and dependence must be accounted for in Monte Carlo method error estimates.

Start with $v^{(0)}>0$ and any coefficient vector. At sweep $m$, draw $a_k^{(m+1)}$ from the multivariate normal distribution above using $v^{(m)}$, then draw $v^{(m+1)}$ from the inverse-gamma distribution using the newly drawn coefficients. For the second step one may generate $1/v$ from a gamma distribution with the same shape and rate equal to the displayed inverse-gamma scale.

Each block update uses an exact conditional distribution, so it preserves the joint posterior distribution. The resulting pairs form a dependent Gibbs sampler sample. With proper priors, positive scale and positive-definite coefficient covariance, these conditionals have full probability support on ${\mathbb{R}}^{k+1}\times (0,\infty )$ and the usual posterior ergodicity applies. A burn-in reduces dependence on initialization; retained sweeps are still serially dependent and Monte Carlo errors should account for that dependence.

For a target probability density function $\pi ({\theta }_1,\dots ,{\theta }_p)$, one systematic sweep of the Gibbs sampler updates coordinates successively:Each update leaves the target invariant: keeping the other coordinates with their target marginal and replacing one coordinate by its conditional distribution reconstructs the target joint probability density function. Therefore composition of the coordinate updates also preserves $\pi$. It defines a Markov chain whose consecutive draws are generally dependent. Under irreducibility, aperiodicity and the usual recurrence conditions it converges to the target law from an arbitrary valid start; a stationary start gives exact target marginals immediately. Invariance alone does not prove these ergodicity conditions, and a systematic full sweep need not itself be reversible.

At temperature $T>0$, target a probability density function proportional to $\exp (f(\theta )/T)$ on a parameter region where it is normalizable, possibly with a specified proper base measure. A Metropolis–Hastings algorithm proposal $q(\theta ,{\theta }^{\prime })$ has acceptance probabilityFor a symmetric proposal, improving moves are accepted and a loss $\Delta$ is accepted with probability $e^{-\Delta /T}$. Run updates while gradually lowering $T$ toward zero. This simulated annealing scheme initially permits exploration and increasingly favors high objective values. Appropriate cooling and mixing conditions are needed for a general global convergence theorem; a finite arbitrary schedule does not guarantee finding a global maximum.

To maximize a positive likelihood function, take $f=\log L$, which has the same maximizers. The resulting likelihood-power annealing target is $L^{1/T}$, and available Gibbs sampler conditional distributions can replace Metropolis proposals.

Conditioned on the parameters, the data augmentation labels are independent. By Bayes' theorem, the conditional probability of component $i$ for observation $j$ isThis is the mixture responsibility. Draw an independent uniform random variable and select $i$ using the cumulative sums of $(r_{j1},\dots ,r_{jk})$. Refresh every label before recomputing the component sufficient statistics for the next Gibbs sampler sweep. Numerically, form the log scores $\log {\omega }_i-\frac{1}{2}\log v_i-(x_j-{\mu }_i{)}^2/(2v_i)$, subtract their maximum and exponentiate before normalization.

Use the data augmentation labels to separate the mixture into groups. Define $n_i={\sum }_j{1}_{\{z_j=i\}}$, $s_i={\sum }_{z_j=i}{x}_j$ and $Q_i({\mu }_i)={\sum }_{z_j=i}(x_j-{\mu }_i{)}^2$. The intended joint posterior distribution is proportional toSumming over all allocation labels recovers the marginal mixture posterior. The Gaussian mixture Gibbs updates with independent priors follow by collecting the terms in each parameter.

For ${\mu }_i$, the conditional exponent isCompleting the square givesFor $v_i$, the conditional density is proportional to $v_i^{-(\alpha +n_i/2)-1}\exp [-(\beta +Q_i({\mu }_i)/2)/v_i]$, henceThe independent prior on ${\mu }_i$ adds neither a $v_i^{-1/2}$ factor nor a ${\mu }_i^2/v_i$ penalty. Finally, combining the allocation counts with the Dirichlet distribution prior givesA Gibbs sampler successively draws from these conditionals, using the most recently updated values. An empty component has $n_i=s_i=Q_i=0$, so its mean and variance conditionals reduce to their proper priors without dividing by an undefined group sample mean.

The marginal Metropolis–Hastings algorithm avoids storing allocation labels, but each component update changes every mixture likelihood term. Cache component contributions, calculate log-likelihoods stably and tune the proposal scale and random-walk Metropolis efficiency during a pilot run; then fix proposal scales or use a justified adaptive scheme. Means and variances may be strongly dependent, making separate random walks inefficient. Weight proposals must respect the simplex, variance proposals must respect positivity, and the displayed acceptance ratios must retain all relevant priors and proposal corrections.

The augmented Gibbs sampler uses elementary conditionals and typically costs order $nk$ for a label sweep, followed by sufficient-statistic updates. It avoids Metropolis rejections in these conditional draws, but does not remove slow mixing: ambiguous allocations can lock means, variances and labels into strongly coupled configurations. Empty components must be updated from their priors, not deleted automatically. The proper inverse-gamma scale $\beta >0$ suppresses arbitrarily small variances even though an unregularized Gaussian-mixture likelihood can be singular.

With exchangeable component priors, label switching creates $k!$ equivalent posterior modes. Ordering component means is an identification constraint, not an innocuous change unless the intended inferential model is made explicit. Predictive density and permutation-invariant summaries avoid arbitrary labelling; component-specific summaries require an appropriate relabelling convention. If the Dirichlet parameters differ between components, exact label symmetry may be broken, but substantial multimodality can remain.

Use dispersed initial states, multiple chains and diagnostics on several scientifically relevant functions. Report effective sample size of a Markov chain and Monte Carlo error estimates, not just a large iteration count. Burn-in can reduce initialization bias but does not prove convergence; thinning cannot by itself cure poor exploration. Closed-form Gibbs updates simplify computation, while reliable inference still requires adequate exploration of the posterior.

For minimization of an energy $E(\vartheta )$, simulated annealing at temperature $t>0$ targets the Boltzmann distribution ${\pi }_t(\vartheta )\propto e^{-E(\vartheta )/t}$ relative to a specified parameter reference measure. With proposal distribution $q({\vartheta }^{\prime }\mid \vartheta )$, accept withAt each temperature run the chain long enough to explore that law, then cool and retain the best state seen. For symmetric proposals every improvement is accepted, while a worsening move is accepted with probability $e^{-\Delta E/t}$. Equilibrium laws concentrate near global minima when their normalizing integrals are finite and the minima are suitably isolated. A particular rapid cooling schedule does not automatically inherit this guarantee.

Likelihood maximization uses energy $E=-\ell$, not $E=\ell$. The source's later wording “$f(p)$ equal to the log-likelihood” conflicts with its preceding minimization convention. The intended maximum-likelihood Boltzmann distribution is proportional to $e^{\ell /t}=L^{1/t}$. With the opposite sign, likelihood maxima would be disfavored; for binomial data with both successes and failures the density often would not even be integrable near the endpoints as $t$ decreases.

For the maximum-likelihood estimates, write $s={\sum }_{j=1}^m{x}_j$, $\bar{x}=s/m$ and $S={\sum }_j(x_j-\bar{x}{)}^2$. In the binomial distribution model with known positive integer $n$, the log-likelihood isDifferentiating gives $s/p-(mn-s)/(1-p)=0$, hence $\hat{p}=s/(mn)$. The second derivative is negative in the interior. If $s=0$ or $s=mn$, the maximum is at the corresponding endpoint. This model requires integer observations in $\{0,\dots ,n\}$; otherwise its likelihood is zero.

For the normal distribution model write $v={\sigma }^2$. Its log-likelihood isFor fixed $v$, the unique maximizing mean is $\bar{x}$. Substituting it and differentiating in $v$ gives $v=S/m$, so, assuming $S>0$,If $S=0$, the normal density likelihood grows without bound as $v\downarrow 0$ with $\mu =\bar{x}$; there is no finite interior maximum or ordinary finite AIC for that fit.

For Gaussian likelihood Gibbs annealing, use $L_N^{1/t}$ relative to $d\mu dv$. Completing the square and identifying an inverse-gamma distribution gives the Gibbs sampler conditionalsIntegrating out $\mu$ contributes a factor proportional to $v^{1/2}$, so the marginal is $v\sim \mathrm{InvGamma}(m/(2t)-3/2,S/(2t))$. The joint target is proper for $0<t<m/3$ when $S>0$; starting the Lebesgue-based annealing above this range would require a proper reference factor or bounded parameter domain. For sufficiently small $t$, its moments satisfyThus equilibrium draws concentrate at $(\bar{x},S/m)$ as $t\downarrow 0$.

There is also a direct convergence proof for the actual successive Gibbs sampler updates here. Choose deterministic $0<t_r\le m/10$ with $t_r\to 0$, start at a finite deterministic $v_0>0$, draw ${\mu }_r$ from the first conditional using $v_{r-1}$, then draw $v_r$ from the second using ${\mu }_r$. Conditional normal moments give $E[({\mu }_r-\bar{x}{)}^2\mid v_{r-1}]=t_r{v}_{r-1}/m$ and $E[({\mu }_r-\bar{x}{)}^4\mid v_{r-1}]=3t_r^2{v}_{r-1}^2/m^2$. The permitted inverse-gamma moments therefore yieldThe coefficient on the preceding first moment is at most $1/6$, and that on the preceding second moment is at most $1/8$. The remaining coefficients are uniformly bounded, so both moment sequences are bounded by iteration. Since $t_r\to 0$, the recursions then give $E[v_r]\to S/m$ and $E[v_r^2]\to (S/m{)}^2$. Also $E[({\mu }_r-\bar{x}{)}^2]=t_{r}E[v_{r-1}]/m\to 0$. HenceThis proves convergence in mean square, and thus in probability, to the normal maximum-likelihood estimates for this explicit cooling/update scheme, rather than merely assuming equilibration.

For binomial likelihood-power annealing, normalizing $p^{s/t}(1-p{)}^{(mn-s)/t}$ on $(0,1)$ givesIts mean is $(s+t)/(mn+2t)$, and its variance isThe mean tends to $s/(mn)$ and the variance tends to zero, proving concentration at the binomial maximum-likelihood estimate, including the endpoint cases. Independent exact draws from these beta laws are a simple annealing implementation within this model.

For model selection, the Akaike information criterion is $2d-2\hat{\ell }$ with $d_B=1$ and $d_N=2$. ThusUse the endpoint convention $0\log 0=0$. For a trans-dimensional annealing for penalized likelihood construction, target the disjoint model spaces with unnormalized densitiesrelative to $dp$ and $d\mu dv$, respectively. The energies are half the unoptimized AIC, $d-\ell$; their joint minimizer is the fitted model with smaller AIC. Keep constants such as $C_B$ and the normal density normalization, which do not cancel across models. The within-model conditional draws are unchanged by these constant penalties.

Here is a complete dimension-changing proposal. Select a binomial-to-normal move with probability $b>0$ and a normal-to-binomial move with probability $d>0$, both fixed; other steps can use the within-model updates. From $B,p$, draw $u\sim N(0,1)$ with density $h(u)=(2\pi {)}^{-1/2}{e}^{-u^2/2}$ and setThis bijection from $(p,u)\in (0,1)\times \mathbb{R}$ onto $({\mu }^{\prime },v^{\prime })\in \mathbb{R}\times (0,\infty )$ has inverse $p=(1+e^{-{\mu }^{\prime }}{)}^{-1}$, $u=\log v^{\prime }$. Its absolute Jacobian determinant is $v^{\prime }/[p(1-p)]$. The Binomial-normal reversible-jump annealing acceptance ratio is thereforeFor a reverse proposal from $N,\mu ,v$, compute $p^{\prime }=(1+e^{-\mu }{)}^{-1}$ and $u^{\prime }=\log v$ deterministically and accept withThese ratios are reciprocal at inverse states and enforce detailed balance on the dimension-matched augmented spaces. They specify the proposal density, model-selection probabilities and Jacobian explicitly. The map is a valid example, not a claim of optimal mixing. Positive-temperature beta draws stay in the open interval even when the limiting estimate is an endpoint. At each temperature ensure adequate model switching; retain the best fitted penalized likelihood encountered. If minima tie, limiting concentration need not choose a unique model.

There is a material measurement qualification. A binomial probability mass and an unrounded normal density use different observation reference measures. Their raw AIC values are not an invariant, scientifically calibrated comparison of the same recorded data. The expressions and jump algorithm above implement the formal density comparison suggested by the question, with its units fixed. For integer-recorded measurements, a compatible normal model instead assigns probabilities $P_{\mu ,v}(x-1/2<Z\le x+1/2)$, suitably adjusting boundaries or truncation if required. Then both models describe the same recorded outcomes relative to counting measure. Use those masses in the normal likelihood and the same reversible-jump formula; its simple normal MLE and conjugate Gibbs updates will generally no longer apply. This observation-model correction cannot be silently combined with the continuous-normal formulas.