# Dealing With Multicollinearity: A Brief Overview and Introduction to Tolerant Methods

This semester I’m taking a Multivariate statistics course taught by Professor Scott Steinschneider in the BEE department at Cornell. I’ve been really enjoying the course thus far and thought I would share some of what we’ve covered in the class with a blog post. The material below on multicollinearity is from Dr. Steinschneider’s class, presented in my own words.

### What is Multicollinearity?

Multicollinearity is the condition where two or more predictor variables in a statistical model are linearly related (Dormann et. al. 2013). The existence of multicollinearity in your data set can result in an increase of the variance of regression coefficients leading to unstable estimation of parameter values. This in turn can lead to erroneous identification of relevant predictors within a regression and detracts from a model’s ability to extrapolate beyond the range of the sample it was constructed with. In this post, I’ll explain how multicollinearity causes problems for linear regression by Ordinary Least Squares (OLS), introduce three metrics for detecting multicollinearity and detail two “Tolerant Methods” for dealing with multicollinearity within a data set.

### How does multicollinearity cause problems in OLS regression?

To illustrate the problems caused by multicollinearity, let’s start with a linear regression: $y=x\beta +\epsilon$

Where: $y=x\beta +\epsilon$ $x = a \hspace{.1 cm} vector \hspace{.1 cm} of \hspace{.1 cm} predictor \hspace{.1 cm} variables$ $\beta = a \hspace{.1 cm} vector \hspace{.1 cm} of \hspace{.1 cm} coefficients$ $\epsilon = a \hspace{.1 cm} vector \hspace{.1 cm} of \hspace{.1 cm} residuals$

The Gauss-Markov theorem states that the Best Linear Unbiased Estimator (BLUE) for each  coefficient can be found using OLS: $\hat{\beta}_{OLS} = (x^Tx)^{-1}x^Ty$

This  estimate will have a variance defined as: $var(\hat{\beta}_{OLS}) =\sigma^2 (x^Tx)^{-1}$

Where: $\sigma^2 = the \hspace{.1 cm} variance\hspace{.1 cm} of \hspace{.1 cm} the\hspace{.1 cm} residuals$

If you dive into the matrix algebra, you will find that the term (xTx) is equal to a matrix with ones on the diagonals and the pairwise Pearson’s correlation coefficients (ρ) on the off-diagonals: $(x^Tx) =\begin{bmatrix} 1 & \rho \\ \rho & 1 \end{bmatrix}$

As the correlation values increase, the values within (xTx)-1 also increase. Even with a low residual variance, multicollinearity can cause large increases in estimator variance. Here are a few examples of the effect of multicollinearity using a hypothetical regression with two predictors: $\rho = .3 \rightarrow (x^Tx)^{-1} =\begin{bmatrix} 1 & \rho \\ \rho & 1 \end{bmatrix}^{-1} = \begin{bmatrix} 1.09 & -0.33 \\ -0.33 & 1.09 \end{bmatrix}$ $\rho = .9 \rightarrow (x^Tx)^{-1} =\begin{bmatrix} 1 & \rho \\ \rho & 1 \end{bmatrix}^{-1} = \begin{bmatrix} 5.26 & -4.73 \\ -5.26 & -4.73 \end{bmatrix}$ $\rho = .999 \rightarrow (x^Tx)^{-1} =\begin{bmatrix} 1 & \rho \\ \rho & 1 \end{bmatrix}^{-1} = \begin{bmatrix} 500.25 & -499.75 \\ -499.75 & 500.25\end{bmatrix}$

So why should you care about the variance of your coefficient estimators? The answer depends on what the purpose of your model is. If your only goal is to obtain an accurate measure of the predictand, the presence of multicollinearity in your predictors might not be such a problem. If, however, you are trying to identify the key predictors that effect the predictand, multicollinearity is a big problem.

OLS estimators with large variances are highly unstable, meaning that if you construct estimators from different data samples you will potentially get wildly different estimates of your coefficient values (Dormann et al. 2013). Large estimator variance also undermines the trustworthiness of hypothesis testing of the significance of coefficients. Recall that the t value used in hypothesis testing for an OLS regression coefficient is a function of the sample standard deviation (the square root of the variance) of the  OLS estimator. $t_{n-2} =\frac{\hat{\beta_j}-0}{s_{\beta_j}}$

An estimator with an inflated standard deviation, $s_{\beta_j}$, will thus yield a lower t value, which could lead to the false rejection of a significant predictor (ie. a type II error). See Ohlemüller et al. (2008) for some examples where hypothesis testing results are undermined by multicollinearity.

### Detecting Multicollinearity within a data set

Now we know how multicollinearity causes problems in our regression, but how can we tell if there is multicollinearity within a data set? There are several commonly used metrics for which basic guidelines have been developed to determine whether multicollinearity is present.

The most basic metric is the pairwise Pearson Correlation Coefficient between predictors, r. Recall from your intro statistics course that the Pearson Correlation Coefficient is a measure of the linear relationship between two variables, defined as: $r_{x_1,x_2}=\frac{cov(x_1,x_2)}{\sigma_x\sigma_y}$

A common rule of thumb is that multicollinearity may be a problem in a data set if any pairwise |r| > 0.7 (Dormann et al. 2013).

Another common metric is known as the Variance Inflation Factor (VIF). This measure is calculated by regressing each predictor on all others being used in the regression. $VIF(\beta_j) = \frac{1}{1-R^2_j}$

Where Rj2 is the R2 value generated by regressing predictor xj on all other predictors. Multicollinearity is thought to be a problem if VIF > 10 for any given predictor (Dormann et al. 2012).

A third metric for detecting multicollinearity in a data set is the Condition Number (CN) of the predictor matrix defined as the square root of the ratio of the largest and smallest eigenvalues in the predictor matrix: $CN=\sqrt{\frac{\lambda_{max}}{\lambda_{min}}}$

CN> 15 indicates the possible presence of multicollinearity, while a CN > 30 indicates serious multicollinearity problems (Dormann et al. 2013).

### Dealing with Multicollinearity using Tolerant Methods

In a statistical sense, there is no way to “fix” multicollinearity. However, methods have been developed to mitigate its effects. Perhaps the most effective way to remedy multicollinearity is to make a priori judgements about the relationship between predictors and remove or consolidate predictors that have known correlations. This is not always possible however, especially when the true functional forms of relationships are not known (which is often why regression is done in the first place). In this section I will explain two “Tolerant Methods” for dealing with multicollinearity.

The purpose of Tolerant Methods is to reduce the sensitivity of regression parameters to multicollinearity. This is accomplished through penalized regression. Since multicollinearity can result in large and opposite signed  estimator values for correlated predictors, a penalty function is imposed to keep the value of predictors below a pre-specified value. $\sum_{j=1}^{p}|\beta|^l \leq c$

Where c is the predetermined value representing model complexity, p is the number of predictors and l is either 1 or 2 depending on the type of tolerant method employed (more on this below).

#### Ridge Regression

Ridge regression uses the L2 norm, or Euclidean distance, to constrain model coefficients (ie. l = 2 in the equation above). The coefficients created using ridge regression are defined as: $\hat{\beta}_{r} = (x^Tx+\lambda I)^{-1}x^Ty$

Ridge regression adds a constant, λ, to the term xTx to construct the estimator. It should be noted that both x and y should be standardized before this estimator is constructed. The Ridge regression coefficient is the result of a constrained version of the ordinary least squares optimization problem. The objective is to minimize the sum of square errors for the regression while meeting the complexity constraint. $\hat{\beta_r} \begin{cases} argmin(\beta) \hspace{.1cm}\sum_{i=1}^{N} \epsilon_i^2 \\ \sum_{j=1}^{p}|\beta_j|^2 \leq c \end{cases}$

To solve the constrained optimization, Lagrange multipliers can be employed. Let z equal the Residual Sum of Squares (RSS) to be minimized: $argmin(\beta) \hspace{.3cm} z= (y-x\beta)^T(y-x\beta)+\lambda(\sum_{i=1}^{N}|\beta_j|^2-c)$

This can be rewritten in terms of the L2 norm of β: $z = (y-x\beta)^T(y-x\beta)+\lambda||\beta||^2_2$

Taking the derivative with respect to β and solving: $0 = \frac{\partial z}{\partial \beta} = -2x^T(y-x\beta)+2\lambda\beta$ $x^Ty = x^Tx\beta+\lambda\beta=(x^Tx+\lambda I)\beta$ $\hat{\beta}_{r} = (x^Tx+\lambda I)^{-1}x^Ty$

Remember that the Gauss-Markov theorem states that the OLS estimate for regression coefficients is the BLUE, so by using ridge regression, we are sacrificing some benefits of OLS estimators in order to constrain estimator variance. Estimators constructed using ridge regression are in fact biased, this can be proven by calculating the expected value of ridge regression coefficients. $E[\hat{\beta_r}]=(I+\lambda(x^Tx)^{-1})\beta \neq \beta$

For a scenario with two predictors, the tradeoff between reduced model complexity and increase bias in the estimators can be visualized graphically by plotting the estimators of the two beta values against each other. The vector of beta values estimated by regression are represented as points on this plot $(\hat{\beta}=[\beta_1, \beta_2])$.  In Figure 1, $\beta_{OLS}$ is plotted in the upper right quadrant and represents estimator that produces the smallest RSS possible for the model. The ellipses centered around  are representations of the increasing RSS resulting from the combination of β1 and β2  values, each RSS is a function of a different lambda value added to the regression.  The circle centered around the origin represents the chosen level of model complexity that is constraining the ridge regression. The ridge estimator is the point where this circle intersects a RSS ellipse. Notice that as the value of c increases, the error introduced into the estimators decreases and vice versa. Figure 1: Geometric Interpretation of a ridge regression estimator. The blue dot indicates the OLS estimate of Beta, ellipses centered around the OLS estimates represent RSS contours for each Beta 1, Beta 2 combination (denoted on here as z from the optimization equation above). The model complexity is constrained by distance c from the origin. The ridge regression estimator of Beta is shown as the red dot, where the RSS contour meets the circle defined by c.

The c value displayed in Figure 1 is only presented to explain the theoretical underpinnings of ridge regression. In practice, c is never specified, rather, a value for λ is chosen a priori to model construction. Lambda is usually chosen through a process known as k-fold cross validation, which is accomplished through the following steps:

1. Partition data set into K separate sets of equal size
2. For each k = 1 …k, fit model with excluding the kth set.
3. Predict for the kth set
4. Calculate the cross validation error (CVerror)for kth set: $CV^{\lambda_0}_k = E[\sum(y-\hat{y})^2]$
5. Repeat for different values of , choose a that minimizes $CV^{\lambda_0} = \frac{1}{k}CV^{\lambda_0}_k$

#### Lasso Regression

Another Tolerant Method for dealing with multicollinearity known as Least Absolute Shrinkage and Selection Operator (LASSO) regression, solves the same constrained optimization problem as ridge regression, but uses the L1 norm rather than the L2 norm as a measure of complexity. $\hat{\beta}_{Lasso} \begin{cases} argmin(\beta) \hspace{.1cm}\sum_{i=1}^{N} \epsilon_i^2 \\ \sum_{j=1}^{p}|\beta_j|^1 \leq c \end{cases}$

LASSO regression can be visualized similarly to ridge regression, but since c is defined by the sum of absolute values of beta, rather than sum of squares, the area it constrains is diamond shaped rather than circular.  Figure 2 shows the selection of the beta estimator from LASSO regression. As you can see, the use of the L1 norm means LASSO regression selects one of the predictors and drops the other (weights it as zero). This has been argued to provide a more interpretable estimators (Tibshirani 1996). Figure 2: Geometric interpretation of Lasso Regression Estimator. The blue dot indicates the OLS estimate of Beta, ellipses centered around the OLS estimate represents RSS contours for each Beta 1, Beta 2 combination (denoted as z from the optimization equation). The mode complexity is constrained by the L1 norm representing model complexity. The Lasso estimator of Beta is shown as the red dot, the location where the RSS contour intersects the diamond defined by c.

### Final thoughts

If you’re creating a model with multiple predictors, it’s important to be cognizant of potential for multicollinearity within your data set. Tolerant methods are only one of many possible remedies for multicollinearity (other notable techniques include data clustering and Principle Component Analysis) but it’s important to remember that no known technique can truly “solve” the problem of multicollinearity. The method chosen to deal with multicollinearity should be chosen on a case to case basis and multiple methods should be employed if possible to help identify the underlying structure within the predictor data set (Dormann et. al. 2013)

### Citations

Dormann, C. F., Elith, J., Bacher, S., Buchmann, C., Carl, G., Carré, G., Marquéz, J. R. G., Gruber, B., Lafourcade, B., Leitão, P. J., Münkemüller, T., McClean, C., Osborne, P. E., Reineking, B., Schröder, B., Skidmore, A. K., Zurell, D. and Lautenbach, S. 2013, “Collinearity: a review of methods to deal with it and a simulation study evaluating their performance.” Ecography, 36: 27–46. doi:10.1111/j.1600-0587.2012.07348.x

Ohlemüller, R. et al. 2008. “The coincidence of climatic and species rarity: high risk to small-range species from climate change.” Biology Letters. 4: 568 – 572.

Tibshirani, Robert 1996. “Regression shrinkage and selection via the lasso.” Journal of the Royal Statistical Society. Series B (Methodological): 267-288.

# Calculating Risk-of-Failures as in the Research Triangle papers (2014-2016) – Part 1

There has been a series of papers (e.g., Palmer and Characklis, 2009; Zeff et al., 2014; Herman et al., 2014) suggesting the use of an approximate risk-of-failure (ROF) metric, as opposed to the more conventional days of supply remaining, for utilities’ managers to decide when to enact not only water use restrictions, but also water transfers between utilities. This approach was expanded to decisions about the best time and in which new infrastructure project a utility should invest (Zeff at al., 2016), as opposed to setting fixed times in the future for either construction or options evaluation. What all these papers have in common is that drought mitigation and infrastructure expansion decisions are triggered when the values of the short and long-term ROFs, respectively, for a given utility exceeds those of pre-set triggers.

For example, the figure below shows that as streamflows (black line, subplot “a”) get lower while demands are maintained (subplot “b”), the combined storage levels of the fictitious utility starts to drop around the month of April (subplot “c”), increasing the utility’s short-term ROF (subplot “d”) until it finally triggers transfers and restrictions (subplot “e”). Despite the triggered restriction and transfers, the utility’s combined storage levels crossed the dashed line in subplot “c”, which denotes the fail criteria (i.e. combined storage levels dropping below 20% of the total capacity). It is beyond the scope of this post to go into the details presented in all of these papers, but even after reading them the readers may be wondering how exactly ROFs are calculated. In this post, I’ll try to show in a graphical and concise manner how short-term ROFs are calculated.

In order to calculate a utility’s ROF for week m, we would run 50 independent simulations (henceforth called ROF simulations) all departing from the system conditions (reservoir storage levels, demand probability density function, etc.) observed in week m, and each using one of 50 years of streamflows time series recorded immediately prior to week m. The utility’s ROF is then calculated as the number of ROF simulations in which the combined storage level of that utility dropped below 20% of the total capacity in at least one week, divided by the number of ROF simulations ran (50). An animation of the process can be seen below. For example, for a water utility who started using ROF triggers on 01/01/2017, this week’s short-term ROF (02/13/2017, or week m=7) would be calculated using the recorded streamflows from weeks 6 through -47 (assuming here a year of 52 weeks, for simplicity) for ROF simulation 1, the streamflows from weeks -48 to -99 for ROF simulation 2, and so on until we reach 50 simulations. However, if the utility is running an optimization or scenario evaluation and wants to calculate the ROF in week 16 (04/10/2017) of a system simulation, ROF simulation 1 would use 10 weeks of synthetically generated streamflows (16 to 7) and 42 weeks of historical records (weeks 6 to -45), simulation 2 would use records for weeks -46 to -97, and so on, as in a 50 years moving window.

In another blog post, I will show how to calculate the long-term ROF and the reasoning behind it.

Works cited

Herman, J. D., H. B. Zeff, P. M. Reed, and G. W. Characklis (2014), Beyond optimality: Multistakeholder robustness tradeoffs for regional water portfolio planning under deep uncertainty, Water Resour. Res., 50, 7692–7713, doi:10.1002/2014WR015338.

Palmer, R., and G. W. Characklis (2009), Reducing the costs of meeting regional water demand through risk-based transfer agreements, J. Environ. Manage., 90(5), 1703–1714.

Zeff, H. B., J. R. Kasprzyk, J. D. Herman, P. M. Reed, and G. W. Characklis (2014), Navigating financial and supply reliability tradeoffs in regional drought management portfolios, Water Resour. Res., 50, 4906–4923, doi:10.1002/2013WR015126.

Zeff, H. B., J. D. Herman, P. M. Reed, and G. W. Characklis (2016), Cooperative drought adaptation: Integrating infrastructure development, conservation, and water transfers into adaptive policy pathways, Water Resour. Res., 52, 7327–7346, doi:10.1002/2016WR018771.

# Synthetic streamflow generation

A recent research focus of our group has been the development and use of synthetic streamflow generators.  There are many tools one might use to generate synthetic streamflows and it may not be obvious which is right for a specific application or what the inherent limitations of each method are.  More fundamentally, it may not be obvious why it is desirable to generate synthetic streamflows in the first place.  This will be the first in a series of blog posts on the synthetic streamflow generators in which I hope to sketch out the various categories of generation methods and their appropriate use as I see it.  In this first post we’ll focus on the motivation and history behind the development of synthetic streamflow generators and broadly categorize them.

### Why should we use synthetic hydrology?

The most obvious reason to use synthetic hydrology is if there is little or no data for your system (see Lamontagne, 2015).  Another obvious reason is if you are trying to evaluate the effect of hydrologic non-stationarity on your system (Herman et al. 2015; Borgomeo et al. 2015).  In that case you could use synthetic methods to generate flows reflecting a shift in hydrologic regime.  But are there other reasons to use synthetic hydrology?

In water resources systems analysis it is common practice to evaluate the efficacy of management or planning strategies by simulating system performance over the historical record, or over some critical period.  In this approach, new strategies are evaluated by asking the question:  How well would we have done with your new strategy?

This may be an appealing approach, especially if some event was particularly traumatic to your system. But is this a robust way of evaluating alternative strategies?  It’s important to remember that any hydrologic record, no matter how long, is only a single realization of a stochastic process.  Importantly, drought and flood events emerge as the result of specific sequences of events, unlikely to be repeated.  Furthermore, there is a 50% chance that the worst flood or drought in an N year record will be exceeded in the next N years.  Is it well advised to tailor our strategies to past circumstances that will likely never be repeated and will as likely as not be exceeded?  As Lettenmaier et al.  reminds us “Little is certain about the future except that it will be unlike the past.”

Even under stationarity and even with long hydrologic records, the use of synthetic streamflow can improve the efficacy of planning and management strategies by exposing them to larger and more diverse flood and drought than those in the record (Loucks et al. 1981; Vogel and Stedinger, 1988; Loucks et al. 2005).  Figure 7.12 from Loucks et al. 2005 shows a typical experimental set-up using synthetic hydrology with a simulation model.  Often our group will wrap an optimization model like Borg around this set up, where the system design/operating policy (bottom of the figure) are the decision variables, and the system performance (right of the figure) are the objective(s). (Loucks et al. 2005)

### What are the types of generators?

Many synthetic streamflow generation techniques have been proposed since the early 1960s.  It can be difficult for a researcher or practitioner to know which method is best suited to the problem at hand.  Thus, we’ll start with a very broad characterization of what is out there, then proceed to some history.

Broadly speaking there are two approaches to generating synthetic hydrology: indirect and direct.  The indirect approach generates streamflow by synthetically generating the forcings to a hydrologic model.  For instance one might generate precipitation and temperature series and input them to a hydrologic model of a basin (e.g. Steinschneider et al. 2014).  In contrast, direct methods use statistical techniques to generate streamflow timeseries directly.

The direct approach is generally easier to apply and more parsimonious because it does not require the selection, calibration, and validation of a separate hydrologic model (Najafi et al. 2011).  On the other hand, the indirect approach may be desirable.  Climate projections from GCMs often include temperature or precipitation changes, but may not describe hydrologic shifts at a resolution or precision that is useful.  In other cases, profound regime shifts may be difficult to represent with statistical models and may require process-driven models, thus necessitating the indirect approach.

Julie’s earlier series focused on indirect approaches, so we’ll focus on the direct approach.  Regardless of the approach many of the methods are same.  In general generator methods can be divided into two categories: parametric and non-parametricParametric methods rely on a hypothesized statistical model of streamflow whose parameters are selected to achieve a desired result (Stedinger and Taylor, 1982a).  In contrast non-parametric methods do not make strong structural assumptions about the processes generating the streamflow, but rather rely on re-sampling from the hydrologic record in some way (Lall, 1995).  Some methods combine parametric and non-parametric techniques, which we’ll refer to as semi-parametric (Herman et al. 2015).

Both parametric and non-parametric methods have advantages and disadvantages.  Parametric methods are often parsimonious, and often have analytical forms that allow easy parameter manipulation to reflect non-stationarity.  However, there can be concern that the underlying statistical models may not reflect the hydrologic reality well (Sharma et al. 1997).  Furthermore, in multi-dimensional, multi-scale problems the proliferation of parameters can make parametric models intractable (Grygier and Stedinger, 1988).  Extensive work has been done to confront both challenges, but they may lead a researcher to adopt a non-parametric method instead.

Because many non-parametric methods ‘re-sample’ flows from a record, realism is not generally a concern, and most re-sampling schemes are computationally straight forward (relatively speaking).  On the other hand, manipulating synthetic flows to reflect non-stationarity may not be as straightforward as a simple parameter change, though methods have been suggested (Herman et al. 2015Borgomeo et al. 2015).  More fundamentally, because non-parametric methods rely so heavily on the data, they require sufficiently long records to ensure there is enough hydrologic variability to sample.  Short records can be a concern for parametric methods as well, though parametric uncertainty can be explicitly considered in those methods (Stedinger and Taylor, 1982b).  Of course, parametric methods also have structural uncertainty that non-parametric models largely avoid by not assuming an explicit statistical model.

In the coming posts we’ll dig into the nuances of the different methods in greater detail.

### A historical perspective

The first use of synthetic flow generation seems to have been by Hazen .  That work attempted to quantify the reliability of a water supply by aggregating the streamflow records of local streams into a 300-year ‘synthetic record.’  Of course the problem with this is that the cross-correlation between concurrent flows rendered the effective record length much less than the nominal 300 years.

Next Barnes  generated 1,000 years of streamflow for a basin in Australia by drawing random flows from a normal distribution with mean and variance equal to the sample estimates from the observed record.  That work was extended by researchers from the Harvard Water Program to account for autocorrelation of monthly flows (Maass et al., 1962; Thomas and Fiering, 1962).  Later work also considered the use of non-normal distributions (Fiering, 1967), and the generation of correlated concurrent flows at multiple sites (Beard, 1965; Matalas, 1967).

Those early methods relied on first-order autoregressive models that regressed flows in the current period on the flows of the previous period (see Loucks et al.’s Figure 7.13  below).  Box and Jenkins  extended those methods to autoregressive models of arbitrary order, moving average models of arbitrary order, and autoregressive-moving average models of arbitrary order.  Those models were the focus of extensive research over the course of the 1970s and 1980s and underpin many of the parametric generators that are widely used in hydrology today (see Salas et al. 1980; Grygier and Stedinger, 1990; Salas, 1993; Loucks et al. 2005). (Loucks et al. 2005)

By the mid-1990s, non-parametric methods began to gain popularity (Lall, 1995).  While much of this work has its roots in earlier work from the 1970s and 1980s (Yakowitz, 1973, 1979, 1985; Schuster and Yakowitz, 1979; Yakowitz and Karlsson, 1987; Karlson and Yakowitz, 1987), improvements in computing and the availability of large data sets meant that by the mid-1990s non-parametric methods were feasible (Lall and Sharma, 1996).  Early examples of non-parametric methods include block bootstrapping (Vogel and Shallcross, 1996), k-nearest neighbor (Lall and Sharma, 1996), and kernel density methods (Sharma et al. 1997).  Since that time extensive research has made improvement to these methods, often by incorporating parametric elements.  For instance, Srinivas and Srinivasan (2001, 2005, and 2006) develop a hybrid autoregressive-block bootstrapping method designed to improve the bias in lagged correlation and to generate flows other than the historical, for multiple sites and multiple seasons.  K-nearest neighbor methods have also been the focus of extensive research (Rajagopalan and Lall, 1999; Harrold et al. 2003; Yates et al. 2003; Sharif and Burn, 2007; Mehortra and Sharma, 2006; Prairie et al. 2006; Lee et al. 2010, Salas and Lee, 2010, Nowak et al., 2010), including recent work by our group  (Giuliani et al. 2014).

Emerging work focuses on stochastic streamflow generation using copulas [Lee and Salas, 2011; Fan et al. 2016], entropy theory bootstrapping [Srivastav and Simonovic, 2014], and wavelets [Kwon et al. 2007; Erkyihun et al., 2016], among other methods.

In the following posts I’ll address different challenges in stochastic generation [e.g. long-term persistence, parametric uncertainty, multi-site generation, seasonality, etc.] and the relative strengths and shortcomings of the various methods for addressing them.

### Works Cited

Barnes, F. B., Storage required for a city water supply, J. Inst. Eng. Australia 26(9), 198-203, 1954.

Beard, L. R., Use of interrelated records to simulate streamflow, J. Hydrol. Div., ASCE 91(HY5), 13-22, 1965.

Borgomeo, E., Farmer, C. L., and Hall, J. W. (2015). “Numerical rivers: A synthetic streamflow generator for water resources vulnerability assessments.” Water Resour. Res., 51(7), 5382–5405.

Y.R. Fan, W.W. Huang, G.H. Huang, Y.P. Li, K. Huang, Z. Li, Hydrologic risk analysis in the Yangtze River basin through coupling Gaussian mixtures into copulas, Advances in Water Resources, Volume 88, February 2016, Pages 170-185.

Fiering, M.B, Streamflow Synthesis, Harvard University Press, Cambridge, Mass., 1967.

Giuliani, M., J. D. Herman, A. Castelletti, and P. Reed (2014), Many-objective reservoir policy identification and refinement to reduce policy inertia and myopia in water management, Water Resour. Res., 50, 3355–3377, doi:10.1002/2013WR014700.

Grygier, J.C., and J.R. Stedinger, Condensed Disaggregation Procedures and Conservation Corrections for Stochastic Hydrology, Water Resour. Res. 24(10), 1574-1584, 1988.

Grygier, J.C., and J.R. Stedinger, SPIGOT Technical Description, Version 2.6, 1990.

Harrold, T. I., Sharma, A., and Sheather, S. J. (2003). “A nonparametric model for stochastic generation of daily rainfall amounts.” Water Resour. Res., 39(12), 1343.

Hazen, A., Storage to be provided in impounding reservoirs for municipal water systems, Trans. Am. Soc. Civ. Eng. 77, 1539, 1914.

Herman, J.D., H.B. Zeff, J.R. Lamontagne, P.M. Reed, and G. Characklis (2016), Synthetic Drought Scenario Generation to Support Bottom-Up Water Supply Vulnerability Assessments, Journal of Water Resources Planning & Management, doi: 10.1061/(ASCE)WR.1943-5452.0000701.

Karlsson, M., and S. Yakowitz, Nearest-Neighbor methods for nonparametric rainfall-runoff forecasting, Water Resour. Res., 23, 1300-1308, 1987.

Kwon, H.-H., U. Lall, and A. F. Khalil (2007), Stochastic simulation model for nonstationary time series using an autoregressive wavelet decomposition: Applications to rainfall and temperature, Water Resour. Res., 43, W05407, doi:10.1029/2006WR005258.

Lall, U., Recent advances in nonparametric function estimation: Hydraulic applications, U.S. Natl. Rep. Int. Union Geod. Geophys. 1991- 1994, Rev. Geophys., 33, 1093, 1995.

Lall, U., and A. Sharma (1996), A nearest neighbor bootstrap for resampling hydrologic time series, Water Resour. Res. 32(3), pp. 679-693.

Lamontagne, J.R. 2015,Representation of Uncertainty and Corridor DP for Hydropower 272 Optimization, PhD edn, Cornell University, Ithaca, NY.

Lee, T., J. D. Salas, and J. Prairie (2010), An enhanced nonparametric streamflow disaggregation model with genetic algorithm, Water Resour. Res., 46, W08545, doi:10.1029/2009WR007761.

Lee, T., and J. Salas (2011), Copula-based stochastic simulation of hydrological data applied to Nile River flows, Hydrol. Res., 42(4), 318–330.

Lettenmaier, D. P., K. M. Latham, R. N. Palmer, J. R. Lund and S. J. Burges, Strategies for coping with drought Part II: Planning techniques for planning and reliability assessment, EPRI P-5201, Final Report Project 2194-1, June 1987.

Loucks, D.P., Stedinger, J.R. & Haith, D.A. 1981, Water Resources Systems Planning and Analysis, 1st edn, Prentice-Hall, Englewood Cliffs, N.J.

Loucks, D.P. et al. 2005, Water Resources Systems Planning and Management: An Introduction to Methods, Models and Applications, UNESCO, Delft, The Netherlands.

Maass, A., M. M. Hufschmidt, R. Dorfman, H. A. Thomas, Jr., S. A. Marglin and G. M. Fair,

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