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A capture-recapture model is a statistical model for repeated observations of identifiable individuals, including those missed on some occasions. A closed-population capture-recapture model estimates a fixed population cardinality; models allowing individuals to enter or leave can instead describe survival and recruitment. The sampling assumptions determine which likelihood function is appropriate.

For the two-sample closed-population capture-recapture model, condition on the sample sizes $M$ and $C$. If the second sample selects $C$ distinct individuals from $N$ by simple random sampling, then the overlap $R$ has the hypergeometric distribution:Here $r$ is an integer and the probability mass function is zero outside these bounds. The binomial coefficients count choices of $r$ marked and $C-r$ unmarked individuals, divided by all equally likely size-$C$ subsets. The hypergeometric distribution has expected valueEquating the observed overlap with this expected value gives the Lincoln-Petersen estimator. The underlying sampling without replacement, probability mass function, and expected value are developed in Raymond P. Guiteras, MIT 14.30 hypergeometric distribution handout, §§1–4 and 8.

For the toy counts $M=20$, $C=25$, and $R=5$, hold the observations fixed and vary the unknown integer $N$. The likelihood function isValues below forty cannot contain the observed set union. Exact comparison of consecutive likelihoods gives two tied maximum-likelihood estimates, $N=99$ and $N=100$. The Lincoln-Petersen estimator returns $100$; the Chapman estimator returns $90$ and is a different rule for estimation.

The vertical scale is $L(N)/{\max }_{n\ge 40}L(n)$. This relative likelihood compares how well candidate statistical parameters explain the same observations; it is not a posterior probability. The displayed range $40\le N\le 300$ is only a plotting window. The likelihood function remains positive for every integer $N\ge 40$ and tends to zero as $N$ increases, so the curve has a continuing right tail. The data do not justify truncating statistical inference at three hundred. The conditional hypergeometric distribution formulation also appears in Louis-Paul Rivest and Mamadou Yauck, Small Sample Estimators for Two-way Capture Recapture Experiments, §3, equation (5).

Zero recaptures require care. With $R=0$ and positive $M,C$, the Lincoln-Petersen estimator has no finite ratio, but the capture-recapture model still gives a valid likelihood function:This likelihood function increases toward one as $N$ grows, so there is no finite maximum-likelihood estimate. For the same sample sizes twenty and twenty-five with zero overlap, forty-five distinct individuals would be observed. Lack of overlap supplies a lower bound but very weak information about large population cardinalities.

In more elaborate capture-recapture models, a first-recapture array $m_{ij}$ counts individuals released at occasion $i$ and next observed at $j$. A probability of survival ${\varphi }_j$ and a probability of detection $P_{j+1}$ contribute a no-observation continuation factor ${\varphi }_j(1-P_{j+1})$. The resulting likelihood function depends on both survival and detection; these probabilities need not be separately identifiable at the last observation interval. Missing-count EM for capture-recapture describes one way to fit such a model, and identifiability remains a separate issue from computing an estimator output.

A closed-population capture-recapture model assumes the same individuals remain throughout the observations: no births, deaths, immigration, or emigration. Its elementary two-sample version additionally requires retained, correctly recognized marks and a second sample chosen by simple random sampling independently of first-sample membership. Mixing and unchanged capture behavior help make this plausible. Equal individual capture probabilities alone do not guarantee that every size-$C$ subset is equally likely; that stronger sampling assumption gives the hypergeometric distribution. See Barker, §§1.1.1 and 1.1.3: closure, marks, and catchability assumptions.

Unequal catchability, mark loss, and population change can invalidate this capture-recapture model. For example, if especially catchable individuals appear repeatedly, overlap can be inflated and the Lincoln-Petersen estimator can underestimate population cardinality. Lost marks can decrease recorded overlap and increase the same ratio estimator, all else equal. Two samples alone cannot establish that the sampling assumptions hold.

A discrete uniform distribution gives every element of a nonempty finite set $S$ the same probability. Write $n=|S|$ for its cardinality. A random variable $X$ with this law has support of a probability distribution $S$ and probability mass functionThe masses sum to $n(1/n)=1$. Conversely, if all $n$ masses equal $p$, their sum is $np=1$, so necessarily $p=1/n$. Thus equal chances determine the whole law.^[ref]

Counting now computes an event's probability: for any set $A$ of possible values,In particular, when $A\subseteq S$, the answer is $|A|/|S|$. The intersection matters: values outside the support of a probability distribution contribute zero. This turns a probability question into a combinatorics question, provided that equal chances are justified by the model.

For an ideal fair six-sided die, $S=\{1,2,3,4,5,6\}$ and each face has probability $1/6$. The event of an even integer is $\{2,4,6\}$, so its probability is $3/6=1/2$; the event $X\ge 5$ has probability $2/6=1/3$. Direct summation gives the expected value $\mathbb{E}X=21/6=7/2$ and the variance $\mathrm{Var}(X)=91/6-(7/2{)}^2=35/12$. The mean $7/2$ is a balance point between possible outcomes. These are exact consequences of stipulated fairness, rather than measured frequencies from physical dice.

Uniformity describes probabilities, not distances between values. For a second constructed model, stipulate equal-chance selection from $S=\{1,2,4,8\}$. Every value has probability $1/4$, despite the unequal gaps $1,2,4$. Here $\mathbb{P}(X\ge 4)=2/4=1/2$, while the expected value and variance areNeither the midpoint of the extreme values, $9/2$, nor a formula using only the cardinality gives these moments. For a general numerical support of a probability distribution, use its actual values:

Choosing an item from a finite list with equal chances is a basic building block of simple random sampling. A bijection that relabels the items preserves equal probabilities, although changing numerical labels can change the expected value and variance. A function that merges labels can destroy uniformity: for the fair-die model, $Y=\min (X,3)$ has masses $1/6,1/6,4/6$ at $1,2,3$, since four faces map to three. Always count the original equally likely outcomes.

An indicator random variable for membership in a set $A\subseteq S$ has a Bernoulli distribution with parameter $|A|/n$; counting that event over $r$ independent repetitions gives a binomial distribution with parameters $r$ and $|A|/n$. Equal chances among individual values need not produce equal chances among aggregated counts.^[ref]

The finite population mean for $N\ge 1$ is the arithmetic mean of all fixed values in the target population. Under simple random sampling, the sample mean is an unbiased estimator of it.

An inclusion probability is the probability that a particular unit belongs to the selected sample. For sample size $n$ in simple random sampling from $N$ labels it is $n/N$. It differs from the selection probability at a particular draw position.

Let ${\mu }_0,{\mu }_1$ be the population means of the numerical vaccination intention score in the two recorded groups. The two-sided hypothesis test comparesThe sample means are $13.48553$ and $13.46565$, giving a difference of sample means of approximately 0.01988 score points, in the order group zero minus group one. The reported group counts are 1,866 and 262; seven observations have a missing group indicator.

Welch's t-test divides that estimated difference by its estimated standard error. Its reported test statistic is $t=0.055032$, with approximately $344.43$ statistical degrees of freedom. This is an unequal-variance comparison; fractional statistical degrees of freedom are appropriate for the Welch approximation. The two groups need not have equal sample sizes or equal population variances.^[ref]

The reported p-value is $0.9561$. Under the specified equal-mean null hypothesis and sampling model, a p-value measures the probability of a result at least as extreme as the observed one. It is not a posterior probability that the null hypothesis is true. There is no rejection at a prespecified 5% significance level. The observed difference of sample means is tiny relative to its estimated standard error.

The 95% confidence interval is approximatelyIt includes zero and differences in either direction. The interval's coverage refers to repeated use of the procedure under its assumptions, rather than to a probability assigned to this fixed unknown parameter or to the spread of individual scores. Nonrejection does not establish that the two population means are identical.

Simple random sampling supports the sampling interpretation, but the score is bounded and its raw distribution need not be a normal distribution. The large group sizes can support an approximate normal sampling distribution of the sample means under regularity conditions. The supplied observational study does not identify a causal explanation of any difference, and the numeric coding of the group labels creates no biological or ordinal interpretation.

A null hypothesis specifies the population comparison to be tested, for example equality of the population means for women and their complement. The alternative hypothesis for a two-sided test is a nonzero difference. A Student's t-test measures an estimated difference in units of its estimated standard error:On the reported calibration, the sampling distribution under the null hypothesis is Student's t-distribution with $1225$ statistical degrees of freedom. Consequently the two-sided test has p-valueWith this many statistical degrees of freedom, the Student t-distribution is close to the standard normal distribution: $|t|=2.5$ exceeds the approximate $5\%$ two-sided critical value $1.96$, but falls short of the $1\%$ value $2.576$. Using the supplied calibration, reject at the $5\%$ significance level, but not at $1\%$. The large statistical degrees of freedom justify this comparison; a value $2.5$ need not be significant for an arbitrary very small number of statistical degrees of freedom.

The p-value is the probability, under the specified null hypothesis and sampling assumptions, of a test statistic at least as discrepant as the observed one. It is not the probability that the null hypothesis is true, nor the probability that the observed difference happened “by chance”. A significance level of $5\%$ controls the long-run type I error rate of the testing procedure when its assumptions hold. Failing to reject at $1\%$ does not prove equality. Statistical significance also does not measure the substantive size or importance of a difference.

The reported sample means place the women's mean $1.4$ scale units above the overall mean. The negative test statistic can reflect the order of subtraction; it does not reverse that observed ordering. An appropriate pre-specified directional one-sided test would have a tail p-value approximately $0.00627$ on the supplied calibration, but one cannot choose that direction after inspecting the results simply to halve a p-value.

There is a design issue behind the numerical calculation: the women are included in the overall sample. Comparing a subgroup mean with its containing sample mean requires accounting for shared observations. If $w$ is the women's sample fraction and ${\overline{x}}_O$ the mean for the remaining respondents, thenThese two estimates are not independent. An ordinary independent-two-sample Student's t-test of women against the entire respondent sample would therefore use an inappropriate standard error. The interpretation above accepts the supplied test statistic and its stated null calibration; a valid analysis of the underlying data should compare disjoint groups, or calculate the shared-sample covariance correctly. The summary alone cannot establish which calculation was used. Simple random sampling supports sampling inference under the stated design, but an observed gender association does not establish a causal effect.

Label batches before consulting the random-number table, so their labels are independent of the test results. A uniformly random permutation of $1,\dots ,16$, restricted to a smaller label set, induces a uniformly random ordering of that set. This provides a simple rejection-and-restriction scheme:

With independent random digits instead, rejection sampling gives the same uniform choices: for A keep only digits $1,\dots ,6$; for C keep only $1,\dots ,5$. For B, use uniform two-digit numbers, retain labels $01,\dots ,15$, and reject repeats within a three-batch sample. Taking residues modulo six or fifteen from a table whose range is not divisible by that number would give unequal probabilities. These are respectively day-stratified simple random sampling, morning/afternoon stratified sampling, and a one-day design using cluster sampling. The printed word “rest” for A is interpreted as “test”, consistently with the surveillance task. If it instead meant leaving one batch untested, choose that omitted batch uniformly and test the other five; this alternative would test 25 batches per week, with inclusion probability $5/6$, and has a different testing budget.

Probability sampling chooses observational units by a specified random mechanism with known inclusion probabilities. This separates design-based representativeness from the convenience of obtaining particular observations. Simple random sampling, stratified sampling and cluster sampling are different designs with different precision and cost properties.

The sampling fraction is the sample size divided by the finite population size. In simple random sampling, it determines how much the finite population correction reduces the variance relative to $S^2/n$.

Sampling without replacement selects distinct labelled units: after a unit is selected it is unavailable for the remaining draws. Distinct labels can carry equal numerical values, so repeated values do not necessarily imply a repeated unit. The rule alone does not specify selection probabilities; a procedure favoring particular units is still sampling without replacement but need not be simple random sampling.

In uniform sequential sampling without replacement, the next label is chosen uniformly among those remaining. For a population of $N$ labels, any ordered sequence of $n$ distinct labels has probability $1/[N(N-1)\cdots (N-n+1)]$. Each size-$n$ subset has $n!$ possible orders, so its probability is ${\left(\genfrac{}{}{0ex}{}{N}{n}\right)}^{-1}$. This gives the fixed-size simple random sampling design. Equivalently, take the first $n$ labels of a uniformly random permutation. Restricting a uniformly random permutation to a specified label set preserves uniformity; biased modulo mappings should not replace uniform selection.

The draw positions have identical marginal probability distributions, but exhibit statistical dependence. If the first label is $i$, the conditional probability that the second label is $i$ is zero, whereas its marginal probability is $1/N$. Thus these draws are not independent and identically distributed when $N>1$ and $n>1$. For a marked subset of labels, counting size-$n$ subsets with binomial coefficients gives the hypergeometric distribution. The resulting negative covariance explains the finite population correction for both counts and sample means.

Here simple random sampling means a random unordered subset $A$ of exactly $n$ distinct labels from $\{1,\dots ,N\}$, with every size-$n$ subset equally likely. The sample size is fixed in advance, $1\le n\le N$, and sampling without replacement prevents duplicate units. This is SRSWOR; some texts also use "simple random sampling" for independent uniform draws with replacement, which is a different design.^[ref]

Every label has inclusion probability $n/N$, but equal inclusion probabilities alone are insufficient. For example, from six labels choose either $\{1,2,3\}$ or $\{4,5,6\}$ with probability $1/2$. Each label is included with probability $1/2$, yet eighteen of the twenty size-three subsets are impossible: this is a two-group cluster sampling design. Unequal probability sampling deliberately changes inclusion probabilities, often to improve precision for a particular estimand, and generally requires an appropriately weighted estimator rather than the plain sample mean.

Consider the original mathematical population with labelled values $(2,4,6,8,10,12)$, and let $n=2$. The fifteen equiprobable subsets, followed by their sample means, areTheir sample means sum to $105$, giving expected value $105/15=7$. Their squared deviations from $7$ sum to $70$, giving variance $70/15=14/3$. This is an enumeration over hypothetical selections from six stipulated numbers, not measured survey data.

With independent uniform sampling with replacement, the $6^2=36$ ordered outcomes each have probability $1/36$. The two orders $(2,4)$ and $(4,2)$ are distinct outcomes, and $(2,2)$ is now possible. Forgetting order does not make the resulting unordered multisets equiprobable: $\{2,4\}$ has probability $2/36$ but $\{2,2\}$ has probability $1/36$. The thirty-six sample means still have expected value $7$, but their squared deviations sum to $210$, giving variance $210/36=35/6$. Avoiding duplicate labels improves precision for this fixed population.