Rich-get-richer (RGR) processes are a type of growth process that describe aspects of many objects and phenomena in the natural and social sciences, e.g., firm size and wealth distributions, citation networks, protein phosphorylation, and lung injury. We will review the derivation of a continuous-time, continuous-space version of the RGR process and then describe connections with survival theory. This article is derived partially from our papers also contains original research. Throughout we will be a little cavalier with our notation and assumptions.

If a system evolves according to an RGR process, this means that larger objects (or objects with more money, more distance, and so on) are more likely to become still larger. It is useful to first examine a process that is not an RGR process. Suppose we have \(N\) objects that are randomly partitioned into \(K\) jars. At each timestep, with equal probability we select one of the \(K\) jars and add an element to that jar. It is pretty obvious by construction that this process does not assign higher probability to jars that have more things in them. In fact, you can work out that this process has an equilibrium distribution (if you change coordinates to a system that grows linearly in time). You should see if you can guess (or, hopefully, derive) what that equilibrium distribution is before reading the rest of this article.

At any rate, a simple modification to this method is as follows: instead of picking one of the jars at random and adding an element to it, we will pick one element at random out of the set of all elements, and then add an element to the jar from which the random element was chosen. This is a subtle change but has significant consequences for the equilibrium distribution. This is also the RGR process described by Udny Yule and Herbert Simon in their seminal papers on this topic.

We would like to know how many groups of size \(k\) there are in the system, where \(k = 1,2,…\). We will suppose that the system initially starts with \(n_0\) elements that are in some initial configuration \(N_k(0)\). In addition, we will add the following twist: at each timestep, with probability \(\rho\) the attachment process as we have just outlined it is in operation. But with probability \(1 - \rho\), a new group forms. We will call \(\rho\) the innovation probability. You can probably imagine that it also has a profound effect on the distribution of group sizes — just think about what the size distribution must look like if \(\rho = 1\) so that every element forms a new group! Moving to continuous time (which is a useful approximation if ticks of the system’s governing clock are close together), the equation describing the number of groups of size \(k\), \(N_k\), is

\[\frac{dN_k}{dt} = (1 - \rho) [\text{Inflow into group $k$} - \text{Outflow from group $k$}] + \rho \delta_{k1}.\]

We will encode the actual mechanics of the attachment process through a function \(r(k)\), which determines how elements are selected for replication. We call this function the attachment kernel. For example, the probability that an element from a group with \(k\) elements is selected when using Simon’s mechanism is \(r(k) = k\), since each element has an equal chance of being picked. Using this attachment kernel, the evolution equation is

\[\frac{dN_k}{dt} = (1 - \rho) [P_{k-1} - P_k] + \rho \delta_{k1},\]

where by \(P_k\) we mean the probability of choosing for replication an element from a group of size \(k\). What is this probability? It is proportional to the number of things in each group, which is \(k\), multiplied by the number of these kinds of groups, \(N_k\). Hence \(P_k = kN_k / \sum_{\ell} \ell N_{\ell} = k N_k / (t + n_0),\) since there must be \(t + n_0\) elements in the system at time \(t\). Thus, the actual functional form of the evolution equation is (finally)

\[\begin{aligned} \frac{dN_k}{dt} &= (1 - \rho) \left[ \frac{(k-1)N_{k-1}}{t + n_0} - \frac{k N_k}{t + n_0} \right] + \rho \delta_{k1}. \end{aligned}\]

This equation actually has a closed-form solution which was communicated to me by Babak Fotouhi. Finding this solution is an interesting problem, but we will not consider it during the rest of this article.

In certain contexts, it makes sense to include spatial distance when describing a system driven by rich-get-richer dynamics. If the distance between groups is small, then the master equation that we just derived is well-approximated by a first-order hyperbolic partial differential equation:

\[\frac{\partial N(x,t)}{\partial t} = -\frac{1 - \rho}{t + n_0} \frac{\partial}{\partial x}[r(x)N(x,t)] + \rho \delta(x - x^*).\]

This equation is defined on the half-open interval \(x\in [x_0, \infty)\) where \(x_0 > 0\). We are being as general as possible and so now are including the arbitrary attachment kernel \(r(x)\) instead of just the Simon kernel \(r(x) = x\). In passage to the continuum spatial limit, the Kronecker delta has become a Dirac delta. When \(x \gg x_0\) and \(\rho \ll 1\), the tail of \(N(x,t)\) is approximately given by the solution to the simpler equation

\[\frac{\partial N(x,t)}{\partial t} = -\frac{1 - \rho}{t + n_0} \frac{\partial}{\partial x}[r(x)N(x,t)].\]

We can solve this equation using separation of variables. We assume that the solution scales as the product of function of time only and of a probability distribution, \(N(x,t) = T(t)p(x)\). The PDE decouples into two equations,

\[(t + n_0)\frac{\dot{T}(t)}{T(t)} = \lambda = -\frac{(1 - \rho)}{p(x)} \frac{d}{dx}[r(x)p(x)].\]

The ODE in time has the solution \(T(t) \propto (t + n_0)^{\lambda}\), while the spatial ODE has \(p(x) = \mathcal{N} \frac{1}{r(x)}\exp [ -\int_{x_0}^x \frac{\lambda}{1 - \rho}\frac{1}{r(s)}ds ]\). The number \(\mathcal{N}\) is a function of the eigenvalue \(\lambda\) and of the innovation rate \(\rho\), though as \(t \rightarrow \infty\) we have \(\lambda \rightarrow 1\) since the mass in the tail of \(N(x,t)\) grows linearly in time. This number is the partition function of the probability distribution \(p(x)\). Defining the operator \(L = -\frac{d}{dx}r(x)\), we see that the spatial ODE is actually an eigenvalue equation for this operator:

\[L[p(x)] = \eta p(x),\]

where \(\eta = \frac{1}{1 - \rho} \geq 1\). 

Survival theory

We will take a brief detour from the discussion of continuum RGR processes to outline some elements of survival theory. Fundamentally, survival theory is just a way of looking at probability theory that asks the question “how much more” instead of “how much”. What we mean by this is as follows: if \(p(x)\) is the pdf of the rv \(X\) and \(P(x)\) is its corresponding cdf, \(P(x) = \int_{x_0}^x p(s)\ ds\), then the survival function of \(X\) is \(P_{\geq}(x) = 1 - P(x) = \int_{x}^{\infty}p(s)\ ds\). This function also goes by the name of complementary cumulative distribution function (ccdf). Usually we think of the rv \(X\) as representing time until some event, like death of a human, failure of a mechanical part, or natural disaster. However, \(X\) might represent some non-temporal random variable, such as revolutions until failure, miles driven until breakdown, or cigarettes smoked until diagnosis. The rv \(X\) is supported on the half-open interval \(x \in [x_0, \infty)\), \(x_0 \geq 0\), since quantities in which we are interested are usually in the future (temporal) or intrinsically non-negative, e.g., distance.

We are interested in the instantaneous probability of failure. That is, given that we have not yet observed a failure event at \(x\), we want to know the probability that a failure event occurs in the interval \([x, x + dx)\):

\[\lambda(x) \equiv p_{X \in [x, x + dx) | X \geq x}(x) = \frac{p(x)}{P_{\geq}(x)}.\]

We call \(\lambda(x)\) the hazard function or hf for short. It is the case that an rv is uniquely characterized by its hf. If we integrate \(\lambda(x)\), we find that

\[\Lambda(x) \equiv \int_{x_0}^x \lambda(s)\ ds = \int_{x_0}^x \frac{p(s)}{P_{\geq}(s)}\ ds = -\int_{x_0}^{x}\frac{dP_{\geq}(s)}{P_{\geq}(s)} = - \log P_{\geq}(x),\]

and hence

\[P(x) =1 - \exp \left[ -\Lambda(x) \right],\]

which completes the proof.

From this result we can derive a few facts about hfs. First, they must not decay to zero too quickly. We must have \(\Lambda(x) \rightarrow \infty\) as \(x \rightarrow \infty\), because \(P(x) \rightarrow 1\) as \(x \rightarrow \infty\) by the definition of a cdf. For example, \(\lambda(x) = \frac{1}{x^2}\) is not the hf of any atomless distribution with support on \([x_0, \infty)\) since \(\Lambda(x) = \int^x \frac{1}{s^2}\ ds = \frac{1}{x}\), which does not diverge as \(x \rightarrow \infty\). We will return to this explicit example later in this article. Second, the shape of the hazard function encodes explicit facts about the failure modes of the system under study. When \(\lambda(x) \rightarrow 0\) monotonically as \(x \rightarrow \infty\), the system exhibits “infant mortality” failure; new parts with low usage are more likely to fail, but with increased useage the parts are less likely to fail. In contrast, when \(\lambda(x) \rightarrow \infty\) as \(x \rightarrow \infty\), older parts are increasingly likely to fail. In real systems, hfs are often modeled as the sum of three distinct functions:

\[\lambda(x) = \lambda_{\text{burn in}}(x) + \lambda_{\text{random}}(x) + \lambda_{\text{wear out}}(x).\]

  • \(\lambda_{\text{burn in}}(x)\) describes burn-in or infant mortality and is a monotone decreasing function;

  • \(\lambda_{\text{random}}(x)\) describes random failure and is constant; and

  • \(\lambda_{\text{wear out}}(x)\) describes wear-out or repetitive usage failure and is a monotone increasing function.

We display pdfs corresponding to each of these types of hf in the figure below.

Hazard functions uniquely characterize pdfs. We display three pdfs
that correspond to three qualitatively different hazard functions:
monotone decreasing, constant, and monotone

Though it is easy to differentiate between the pdf corresponding to the monotone increasing hf \(\lambda_{\text{wear out}} = cx\) and the others, it is not as easy to visually distinguish between the pdfs corresponding to \(\lambda_{\text{random}} = c\) and \(\lambda_{\text{burn in}} = c/x\). We therefore display these same pdfs in doubly logarithmic space below.

 We display the same pdfs as above but in doubly-logarithmic space.

RGR processes and survival theory

Now that we have a brief common background on the basics of survival theory, we will discuss the connections between the stationary distribution (again, stationary in the linearly-growing coordinate system) of the RGR process and some quantities that we have derived above. We present these connections here because we have never seen them referenced in any paper on the subject of either RGR (or preferential attachment) processes or in the survival theory literature; if you are familiar with such references, we would appreciate your sharing them with us!

We showed that the eigenvalue equation describing the stationary distribution, \(L[p(x)] = \eta p(x)\), had the solution

\[p(x) = \mathcal{N} \frac{1}{r(x)}\exp [ -\int_{x_0}^x \eta \frac{1}{r(s)}ds ].\]

Integrating this equation, we see that

\[P_{\geq}(x) \propto \exp[ - \int_{x_0}^x \eta \frac{1}{r(s)}ds],\] and hence that the hf for the rv \(X\), \(\lambda(x)\), must be \(\lambda(x) = \eta /r(x)\). Using this result, we can immediately observe several interesting facts:

  • No RGR process that is actually well-defined over the entire semi-infinite interval \([x_0, \infty)\) can have an attachment kernel that scales as any monomial \(r(x) \propto x^\alpha\) with \(\alpha > 1\). If this were possible, then \(\Lambda(x) \propto \int \frac{1}{r(x)} dx \propto x^{-(\alpha - 1)}\), which decays to zero as \(x \rightarrow \infty\). Hence \(\Lambda(x)\) is not a cumulative hazard function, which is a contradiction.

  • What this actually means is that, if a measured \(r(x)\) of an RGR process does grow like \(x^{\alpha}\) with \(\alpha > 1\), at least one of the two following statements is true:

    • The system described by the RGR process has an upper bound on the rv \(X\).

    • The true \(r(x)\) eventually scales more slowly than \(x^{\alpha}\) for large enough values of \(x\).

    If realized values of \(X\) are actually bounded above by \(y\), then \(r(x)\) can grow as fast as it wants for \(x < y\) since \(r(x) = 0\) for all \(x \geq y\). The estimator of \(r(x)\) constructed from \(N\) observations \(X_1,…,X_N\) can certainly scale as \(x^{\alpha}\) with \(\alpha > 1\) since by definition they will all be less than or equal to \(y\). On the other hand, if the system actually does not have an upper bound (for example, accumulation of nominal wealth by economic agents), then the true RGR kernel must surely eventually grow more slowly than \(x^{\alpha}\) with \(\alpha > 1\).

  • Every existing method for estimating the hazard function from data can now be used to measure the RGR kernel from data. We will consider a few common nonparametric methods for estimating the hazard function presently and demonstrate their applicability in estimating \(r(x)\).

  • The mechanism behind the RGR process (or, more generally, behind the continuum growth process with \(r(x)\) not necessarily exhibiting rich-get-richer behavior) can be seen to apply to many scenarios to which survival theory is typically applied.

We will demonstrate how some existing methods of estimating survival or hazard functions can be used to estimate preferential attachment kernels. Even though the formal hf - RGR connection is technically valid only in the continuum asymptotic limit, we will use this correspondence in the original discrete system context.

Throughout we will suppose that \(X_1 < X_2 <…< X_N \) are observed marks at which at least one event (failure, death, wealth level, etc.) happened. By “marks” we mean times, distances, revolutions, or whatever else is being measured as a quantity until failure random variable. At the \(n\)-th mark, \(d_n\geq 0\) failure events occur. (If time is continuous, \(d_n =1 \) almost surely.) A total of \(K_n\) draws from the RGR process have not experienced failure by mark \(X_n\).

Kaplan-Meier estimator

This is a nonparametric method to estimate the sf of the rv \(X\) and is given by \(\hat{P}{\geq}(x) = \prod{ n: X_n < x } (1 - d_n/K_n).\) (For some reason the \(\LaTeX\) of the above formula won’t render. Please inspect the source of this document and see if you can figure out why; we sure can’t.) The cumulative hazard function can therefore be estimated by

\[\hat{\Lambda}(x) = -\sum_{ n: X_n < x } \log (1 - d_n/K_n).\]

Being somewhat cavalier about our definition of the derivative (we can make this as formal as we please by an appeal to absolute continuity of measure and the Radon Nikodym theorem, if we so cared), the RGR kernel is then approximated by

\[\hat{r}(x) = -\eta\left[ \frac{d}{dx} \sum_{ n: X_n < x } \log (1 - d_n/K_n) \right]^{-1}.\]

Nelson - Aalen estimator

Looking at the cumulative hazard function estimator version of the Kaplan - Meier estimator, \(\hat{\Lambda}(x) = -\sum_{ n: X_n < x } \log (1 - d_n/K_n)\), we notice that \(\log(1 - y) \approx -y\) for small \(y\). Using this substitution, we arrive at the alternative cumulative hazard estimator

\[\hat{\Lambda}(x) = \sum_{ n: X_n < x } d_n/K_n.\]

Applying a similar argument, we have

\[\hat{r}(x) = \eta\left[ \frac{d}{dx} \sum_{ n: X_n < x } d_n/K_n \right]^{-1}\]

Let’s take a look at these hf estimators applied to some manufactured datasets. We will draw values from three distributions, one calm (folded normal) and two leptokurtic (fatigue-life and pareto).

 Kaplan-Meier and Nelson-Aalen hazard function estimators applied to
draws from a folded normal distribution.

 Kaplan-Meier and Nelson-Aalen hazard function estimators applied to
draws from a fatigue life distribution.

 Kaplan-Meier and Nelson-Aalen hazard function estimators applied to
draws from a pareto distribution.

Already we can see the problem: we are 1) taking the derivative of a noisy process and 2) subsequently taking the reciprocal of that derivative. These operations are guaranteed to add noise — potentially a lot of noise — to the estimated quantity. And that’s exactly what’s happening here: the hf estimators of the leptokurtic distributions are fine for low values of \(x\) but are dominated by noise in higher ranges of \(x\). It becomes essentially impossible to discern the mean function of the estimated hf.

Multi-stage estimation

We’ll find a different way of estimating \(r(x)\). Namely, we’ll compute an estimator of \(\Lambda(x)\) and then apply some method to recover a smooth function approximation to the noisy estimator. Differentiating and taking the reciprocal of this smooth function should work a little better than the approach we tried above.

We will illustrate this approach using some simulated rich-get-richer processes. We simulate four RGR processes in discrete time for \(N_t = 2\times 10^4\) timesteps with an innovation rate of \(\rho = 0.1\) and four different attachment kernels, \(r(x) \in \{ 1, \log(x + 1), x^{2/3}, x \}\). These attachment kernels lead to qualitatively different functional forms of the failure pdf \(p(x)\), as you can verify through substitution and integration. (If you do attempt to do this, you’ll find yourself computing the integral \(\int_{x_0}^x \frac{ds}{\log(s + 1)}\), which isn’t expressible analytically. This is called, creatively, the logarithmic integral function. As \(x\) grows large, this function increasingly behaves like \(x/\log x\).) We simulated these processes using, which you can use yourself to run other simulations. We then calculated the Nelson - Aalen estimator of the cumulative hazard function for each process, which we display below.

 Estimated cumulative hazard functions using the Nelson - Aalen
estimator. These hazard functions correspond to the displayed RGR kernel

In order to avoid assuming a functional form for the RGR kernel (remember, we’re pretending that we don’t actually know how the data were generated, only that we hypothesize that an RGR mechanism is at work) we will use a nonparametric method to infer smooth functions with these estimators. We model the estimated cumulative hazard functions using Gaussian processes with RBF kernel functions and display the results of this estimation in the below figure.

 Gaussian process-smoothed cumulative hazard functions.

We display the mean function of the Gaussian process in solid curves and the standard deviation of the process in the shaded regions. To be clear, the standard deviation is omly the standard deviation of the latent cumulative hazard function, not the standard deviation of any noise added to the latent cumulative hazard function. Since draws from Gaussian processes with RBF kernels are infinitely differentiable, we shouldn’t (emphasis on shouldn’t) have much of an issue with differentiating and taking the reciprocal of these functions. One more techncial note: since we actually fit the Gaussian process in log - linear space, we have to compute the RGR kernels using \(\frac{d\Lambda(x)}{dx} = \frac{d\Lambda(\log_{10}x)}{d\log_{10}x} \frac{1}{\log(10) x}\). With this in mind, we compute the estimated RGR kernels and display them below along with the true \(r(x)\). We plot the estimated kernels in solid curves and true kernels in dashed curves.

Estimated \(r(x)\) using Gaussian processes. Not terrible for
literally no parameter tuning or

Not terrible for literally no (hyper)parameter tuning or optimizations. There are definitely some issues, though. For example, in the non-rich-get-richer (or rich just stay pretty rich) case where \(r(x) \propto 1\), the estimated kernel oscillates a little around the true function — if we designed our estimation procedure a little better, we could probably eliminate these oscillations.

In short, we have shown a way to estimate the rich-get-richer kernel function that is statistically well-founded and nonparametric. There’s a fair amount of work to do on tuning this method to be more robust, but this methodology could be a useful complement to existing algorithms.


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