# Why it works

# 1 Introduction

## 1.1 What are we going to do in this vignette

In this vignette, we’ll explain the statistical model that the `primarycensored` package is based on. We’ll cover the following key points:

1. Introduction to the censored data problems in time to event analysis.
2. Discuss the relevant issues from censoring in epidemiological data.
3. Introduce the statistical model used in `primarycensored`.
4. Distributions where we can derive the censored delay distribution analytically.

If you are new to the package, we recommend that you start with the `vignette("primarycensored")` vignette.

# 2 Censoring and right truncation problems in time to event analysis

Time-to-event analysis, also known as survival analysis, concerns estimating the distribution of delay times between events. A distinctive feature of the field are the methodological techniques used to deal with the missing data problems common in data sets of delay times[1]. In `primarycensored` we focus on two particular missing data problems:

- Interval censoring. The primary (start) time and/or the secondary (end) time of the delay are unobserved but both are known to be within intervals.
- Right truncation. Truncated delay data items are only reported if their secondary events are less than a known time, for example the current data collection time.

In statistical epidemiology, these missing data problems occur frequently in both data analysis and theoretical modelling. For a more detailed description of these problems in an epidemiological context see[2,3].

For data analysis: In epidemiology, events are commonly reported as occurring on a particular day or week. This means that both events that define a delay are interval censored. In an emerging outbreak, datasets can also be incompletely observed (right truncation). `primarycensored` provides statistically rigorous methods to account for this double censoring and truncation.

For theoretical modelling: Discrete-time models (e.g., `EpiNow2`, `EpiEstim`) implicitly treat both events as interval censored at the model’s timestep. `primarycensored` provides a principled approach to discretising continuous distributions that accounts for this.

# 3 Statistical model used in `primarycensored`

As described in Getting started with `primarycensored`, `primarycensored` focuses on a subset of methods from time to event analysis that address data missingness problems commonly found in epidemiological datasets. We present the statistical problem as a double interval censoring problem, where both the primary event time and the secondary event times are interval censored. We can recover single interval censoring problems by reducing one of the intervals to a point. In particular, all of the methods in `primarycensored` start by assuming no secondary event censoring and solving the primary censoring problem. This is the reason for the packages name despite the most common use case being when there is double censoring. A key assumption we make throughout is that the censoring window for events are known and independent of the event time within the censoring interval. This is known as non-informative censoring.

The target for inference is the distribution of the delay time between the primary and secondary events. We assume that the delay time is a random variable \(T = S - P_{u}\) with distribution function \(F_T(t) = Pr(T < t)\) and density function \(f_T(t)\). In this treatment we assume that the delay time is shift-invariant, that is, the distribution of the delay time is the same regardless of the primary event time.

The (unconditional) primary event time is a random variable \(P_{u}\) with distribution function \(F_{P_{u}}(t)\) and density function \(f_{P_{u}}(t)\). The secondary event time is a random variable \(S\), but in this treatment we construct the secondary event time from the primary time and delay, therefore the marginal distribution of \(S\) is not considered.

The censoring window for each event is the interval within which each event is known to have occurred, respectively, \(P \in [t_P, t_P + w_P]\) and \(S \in [t_S, t_S + w_S]\). The lengths of the censoring windows are \(w_P\) and \(w_S\) respectively. The precise event times within their windows are unobserved. Note that since the primary event time is known up to the censoring window, we are predominantly interested in the conditional primary time \(P = P_{u} | \{ P_{u} \in [t_P, t_P + w_P]\}\) which has density function:

\[ \begin{aligned} f_{P}(p) &= {f_{P_{u}}(p) \over F_{P_{u}}(t_P + w_P) - F_{P_{u}}(t_P)}, \qquad &p \in [t_P, t_P + w_P],\\ f_{P}(p) &= 0, \qquad &\text{otherwise}. \end{aligned} \]

In this note, we measure the censored delay time \(T_c\) between the primary and secondary event windows from endpoint to endpoint: \(T_c = t_S + w_S - (t_P + w_P)\). Note that in the generative model for delays from “Getting started” the truncated delay \(t_{\text{valid}}\) is measured from startpoint to startpoint of event windows.

In our treatment below we focus on the survival function of the time after the end of the primary window to the secondary event time which we denote \(S_{+}\). We then use this to derive the distribution of the censored delay time \(T_c\). This is equivalent to, but differs in mathematical approach from other treatments of the censoring problems in epidemiology, such as[2], see section Connections to other approaches for details.

## 3.1 Censored delay time distribution

In this section, we explain how to derive the distribution of the censored delay time \(T_c\) from the distribution of the delay time \(T\) and the condition distribution of the primary event time \(P\).

### 3.1.1 Survival function of time from the end of the primary censoring window to the secondary event time

When reasoning about the distribution of the censored delay time \(T_c\), it is useful to consider the time from the end (right) point of the primary censoring interval to the secondary time as a random variable,

\[ S_+ = S - (t_P + w_P) = T - ((t_P + w_P) - P) = T - C_P. \]

Where \(T\) is the delay distribution of interest and \(C_P = (t_P + w_P) - P\) is interval between the end (right) point of the primary censoring window and the primary event time; note that by definition \(C_P\) is not observed but we can relate its distribution to the distribution of \(P\): \(F_{C_P}(p) = Pr(C_P < p) = Pr(P > t_P + w_P - p)\).

With non-informative censoring, it is possible to derive the upper distribution function of \(S_+\), or survival function of \(S_+\), from the distribution of \(T\) and the distribution of \(C_P\).

\[ \begin{equation} \begin{split} Q_{S_+}(t) &= Pr(S_+ > t) \\ &= Pr(T > C_P + t) \\ &= \mathbb{E}_{C_P} \Big[Q_T(t + C_P)\Big] \\ &= \int_0^{w_P} Q_T(t + p) f_{C_P}(p) dp. \end{split} \end{equation} \]

Using integration by parts gives: \[ Q_{S_+}(t) = Q_T(t + w_P) + \int_0^{w_P} f_T(t+p) F_{C_P}(p) dp. \tag{3.1} \]

Where we have used that \(Q^{'}_{T} = - f_T\), \(Q_T\) is the survival function of the actual delay distribution of interest and \(w_P\) is the length of the primary censoring window.

Equation (3.1) is the key equation in this note and is used to derive the distribution of the censored delay time \(T_c\). It has the interpretation that the probability that the secondary event time is greater than \(t\) after the end of the primary censoring window is the sum of two disjoint event probabilities:

1. The probability that the actual delay time \(T\) is greater than \(t + w_P\).
2. The probability that the actual delay time \(T\) is between \(t\) and \(t + w_P\), and the primary event time \(P\) occurred sufficiently close to the end of the primary censoring window that the secondary even occurred more than time \(t\) after the end of the primary window.

Note that in “Getting started” the target for numerical quadrature is the cumulative distribution function of the sum of the primary time within the primary censoring window and the delay time.

### 3.1.2 CDF form of \(S_+\) for uniform primary

For uniform primary, equation (3.1) can be rearranged into a CDF-direct form that is often more convenient for computation. Applying \(F_{S_+} = 1 - Q_{S_+}\) and \(F_T = 1 - Q_T\) and integrating the partial expectation by parts, \(\int_t^{t+w_P} f_T(z)\, z\, dz = [z F_T(z)]_t^{t+w_P} - \int_t^{t+w_P} F_T(z)\, dz\), gives

\[ F_{S_+}(d) = \frac{1}{w_P}\Big[ d\, F_T(d) - q\, F_T(q) - \int_q^d F_T(z)\, dz \Big], \]

where \(q = \max(d - w_P, 0)\). Substituting the partial expectation in the Gamma, Log-Normal and Weibull cases gives the equivalent “distribution mean” form

\[ F_{S_+}(d) = \frac{1}{w_P}\Big[ d\, F_T(d;\cdot) - q\, F_T(q;\cdot) - E\big( \tilde F(d) - \tilde F(q) \big) \Big], \]

where \(E\) is the distribution mean and \(\tilde F\) is a shifted or related CDF specific to each distribution. See `vignette("analytic-solutions")` for the per-distribution expressions.

### 3.1.3 Probability of secondary event time within a secondary censoring window

Having constructed the survival function of \(S_+\) with equation (3.1), using numerical quadrature or in some other way, we can calculate the probability mass of a secondary event time falling within a observed secondary censoring window of length \(w_S\) that begins at time \(n - w_S\) after the primary censoring window. This is the probability that the censored delay time \(T_c\) is \(n\).

This gives the censored delay time probability by integrating over censored values:

\[ Pr(S_+ \in [n - w_S, n)) = Q_{S_+}(n-w_S) - Q_{S_+}(n). \tag{3.2} \]

Note that the censored secondary event time can also occur within the primary censoring window. This happens with probability, \[ Q_{S_+}(-w_P) - Q_{S_+}(0) = 1 - Q_T(w_P) - \int_0^{w_P} f_T(p) F_{C_P}(p) dp = Pr(T< C). \]

# 4 Connections to other approaches

In this section, we discuss how the approach taken in `primarycensored` relates to some other approaches to the censored data problems in time to event analysis.

## 4.1 Connection to Park et al 2024

Using the notation from the methods overview given in Park et al[2], we write the conditional probability of the secondary event time \(S\in (S_L,S_R)\) given the primary event time \(P \in (P_L,P_R)\) as:

\[ \begin{aligned} \mathrm{Pr}(S_L < S < S_R | P_L < P < P_R) &= \frac{\mathrm{Pr}(P_L < P < P_R, S_L < S < S_R)}{\mathrm{Pr}(P_L < P < P_R)} \\ &= \frac{\int_{P_L}^{P_R} \int_{S_L}^{S_R} g_P(x) f_x(y-x) dy dx}{\int_{P_L}^{P_R} g_P(x) dx}\\ &= \int_{P_L}^{P_R} \int_{S_L}^{S_R} g_P(x|P_L, P_R) f_x(y-x)dy dx \end{aligned} \]

In this note, we assume that the forward distribution doesn’t vary over time (such that \(f_x = f\)), then

\[ \int_{P_L}^{P_R} \int_{S_L}^{S_R} g_P(x|P_L, P_R) f_x(y-x)dy dx = \int_{P_L}^{P_R} g_P(x|P_L, P_R) \big[F(S_R - x) - F(S_L - x)\big] dx \]

Then, by using integration by parts, we get:

\[ \begin{split} \int_{P_L}^{P_R} g_P(x|P_L, P_R) \big[F(S_R - x) - F(S_L - x)\big] dx &= F(S_R - P_R) - F(S_L - P_R) \\ & - \int_{P_L}^{P_R} G_P(x|P_L, P_R) \big[f(S_L - x) - f(S_R - x)\big] dx \end{split} \tag{4.1} \] Where we have used that \(\partial_x F(S_R - x) = - f(S_R - x)\) and \(\partial_x F(S_L - x) = - f(S_L - x)\).

We can now compare this to equation (3.2) by considering the following transformations:

- \(P_L = -w_P\) and \(P_R = 0\), this in this note we are treating the endpoint of the primary censoring window as the origin.
- \(S_L = n-w_S\) and \(S_R = n\), that is that we are interested in the probability of the secondary event time falling within the secondary censoring window \([n, n+ w_S)\).

Then equation (4.1) becomes:

\[ \begin{aligned} \mathrm{Pr}(S_L < S < S_R | P_L < P < P_R) &= F(n) - F(n-w_S) - \int_{-w_P}^{0} G_P(x|-w_P, 0) \big[f(n - w_S - x) - f(n - x)\big] dx \end{aligned} \] Making the transformation \(x = -p\), and rewriting in the notation of this note gives: \[ \begin{aligned} &= F(n) - F(n-w_S) + \int_{w_P}^{0} G_P(-p|-w_P, 0) \big[f_T(n - w_S + p) - f_T(n +p)\big] dp \\ &= F(n) - F(n-w_S) + \int_{0}^{w_P} G_P(-p|-w_P, 0) \big[f_T(n + p) - f_T(n - w_S +p)\big] dp\\ &= F(n) - F(n-w_S) + \int_{0}^{w_P} (1 - F_{C_P}(p)) \big[f_T(n + p) - f_T(n - w_S +p)\big] dp\\ &= F(n + w_P) - F(n-w_S + w_P) + \int_{0}^{w_P} [f_T(n + p - w_S) - f_T(n + p)] F_{C_P}(p) dp\\ &= Q_T(n-w_S + w_P) - Q_T(n + w_P) + \int_{0}^{w_P} [f_T(n + p - w_S) - f_T(n + p)] F_{C_P}(p) dp \\ &= Q_{S_+}(n-w_S) - Q_{S_+}(n ). \end{aligned} \] which is same as equation (3.2).

In this derivation, we have used that \(G_P(x|-w_P, 0)\) is the distribution function from the time from the start of the primary interval until primary event time, and \(F_{C_P}\) is the distribution function of the time until the end of the primary event window from the primary event time. Therefore, \(G_P(-p|-w_P, 0) = Pr(P < -p | P \in (-w_P, 0)) = 1 - Pr(C_P \leq p) = 1 - F_{C_P}(p)\).

# 5 Learning more

- For more mathematical background on the analytic solutions see the `vignette("analytic-solutions")`.
- For a more introductory explanation of the primary event censored distribution see the `vignette("primarycensored")`.

# References

1.
Leung, K.-M., Elashoff, R. M., & Afifi, A. A. (1997). Censoring issues in survival analysis. Annual Review of Public Health, 18(1), 83–104.
2.
Park, S. W., Akhmetzhanov, A. R., Charniga, K., Cori, A., Davies, N. G., Dushoff, J., Funk, S., Gostic, K., Grenfell, B., Linton, N. M., Lipsitch, M., Lison, A., Overton, C. E., Ward, T., & Abbott, S. (2024). Estimating epidemiological delay distributions for infectious diseases. bioRxiv. https://doi.org/10.1101/2024.01.12.24301247
3.
Charniga, K., Park, S. W., Akhmetzhanov, A. R., Cori, A., Dushoff, J., Funk, S., Gostic, K. M., Linton, N. M., Lison, A., Overton, C. E., Pulliam, J. R. C., Ward, T., Cauchemez, S., & Abbott, S. (2024). Best practices for estimating and reporting epidemiological delay distributions of infectious diseases. PLoS Comput. Biol., 20(10), e1012520. https://doi.org/10.1371/journal.pcbi.1012520